Recent Advances in Statistical Methods: Proceedings of Statistics 2001 Canada : The 4th Conference in Applied Statistics Montreal, Canada 6-8 July 2001

I TW{khn~^-y) dX{v) m

-

"

Jit

\2

rkh"

u

2^"" ¥>(&) 19 12 ^fc=l

From this result and by checking that fM •M-Mn) does, we obtain

1

^>'2{y)/f{y) d\(y) tends to zero as

f^VfjU^

l

f/oW 2 ,w , n

Acknowledgments We thank the referees for interesting questions and remarks.

41

References 1. A. R. Barron, in Proc. IEEE Int. Symp Inform. Theory, Kobe, Japan, (1988). 2. A. R. Barron, L. Gyorfi, and E. C. van der Meulen, IEEE Trans, on Information Theory 38, 1437 (1992). 3. A. R. Barron and C. Sheu, Ann. Statist. 19, 1347 (1991). 4. A. Berlinet G. and Biau, Iterated modified histograms (Submitted for publication, 2001). 5. A. Berlinet, L. Gyorfi snd E. C. van der Meulen, Publications de I'ISUP. 41, 3 (1997). 6. A. Berlinet, I. Vajda and E. C. van der Meulen, IEEE Trans, on Inform. Theory 44, 999 (1998). 7. L. Gyorfi, F. Liese, I. Vajda and E. C. van der Meulen, Statistics 32, 31 (1998). 8. L. Gyorfi, and E. C. van der Meulen, in Transactions of the Twelfth Prague Conference on Information Theory, Statistical Decision Functions and Random Processes, 88 (1994). 9. P. Hall, Ann. Statist. 15, 1491 (1987).

T H E A S Y M P T O T I C D I S T R I B U T I O N OF SPACINGS OF O R D E R STATISTICS M I K E L I S G. B I C K I S Mathematical Sciences Group, Department of Computer Science, Saskatchewan, 106 Wiggins Road, Saskatoon, SK SIN 5E6, E-Mail: [email protected]

University Canada

of

It is well known that for an i.i.d. sample from a uniform distribution, the spacings between the order statistics are asymptotically distributed like i.i.d. exponential random variables. Although the spacings between order statistics from an arbitrary distribution are not in general identically distributed, one can still consider the distribution function to which the empirical cdf of such spacings from an i.i.d. sample would converge as the sample size goes to infinity. The limit of this empirical cdf is called the asymptotic spacing distribution. If X is a random variable with density / , then the asymptotic spacing distribution is 1 — L, where L is the Laplace transform of the distribution of f(X). Although the explicit form for L is not tractable in general, one can use Tauberian theorems to relate the tail behaviour of L to the order of contact of / to the z-axis: The more gradual the approach of / to the axis, the heavier the tail of the asymptotic spacing distribution.

1

Introduction

Imagine a large number of marathon runners crossing a finish line, and consider the set of time intervals between their arrivals. What would be the distribution of the lengths of these intervals? The question is ill-posed, since in general these lengths would not have the same distribution. Nonetheless, one can calculate the empirical distribution of such a collection and ask the question "Is there a probability distribution to which this empirical distribution converges as the sample size goes to infinity?" It was noted by Katz l that the empirical distribution appears to follow a "power law". More precisely, one could say that the tail of the distribution function behaves like a negative power of its argument (although, of course, integrability requirements prohibit such behaviour extending to the origin). This observation led Katz to speculate the spacings in an ordered Normal sample would have such a "power law" distribution. In this paper, we study the relationship between a parent distribution and the asymptotic distribution of spacings between its order statistics. We characterize the distributions for which the spacings distribution has a negative power tail, and discover that these do not, in fact, include the Normal distribution.

42

43

2

Empirical distribution of spacings

Let Xi,...,

Xn be i.i.d. from a distribution F, and define Yi = X(i+1) --X"(i),

i=

l,...,n-l

to be the spacings between consecutive order statistics. We will assume in what follows that F is absolutely continuous. It is well known that if F is the uniform distribution on the unit interval then nYi, i = 1 , . . . ,n — 1 are asymptotically i.i.d. exponential, and thus it makes sense to talk about the asymptotic distribution of the spacings. In general, however, the Yi's will not be identically distributed, even asymptotically, so care is needed in defining what is meant by "the" asymptotic distribution. Let 1

Sn(t) =

n—1

- ^ / ( ^ W )

(1)

n — 1 *-^ i=\

be the empirical survivor function of the scaled spacings, where /(t )00 ) (') represents the indicator function of the interval (t, oo). (Formulas look simpler in this context if we deal with survivor functions rather than cumulative distribution functions. It should cause no confusion if we refer to a distribution by its survivor function rather than its cdf.) Suppose that there is a distribution S such that with probability 1, Sn converges weakly to S. Then we will call S the asymptotic spacings distribution (ASD) of the distribution F. Pyke 2 considered the limit of (1), and presented an elegant proof that oo

/

e-tf(x)f(x)dx

in probability,

-oo

where / = F' is the probability density function of the original sample. He added, without proof, the stronger statement that the convergence is, in fact, uniform with probability 1, referring to a paper by Blum and Weiss 3 . Monte-Carlo investigation led the present author to speculate that the empirical distribution of spacings is, moreover, asymptotically equivalent to the empirical distribution of an appropriate i.i.d. sample, as suggested by the following heuristic argument, under the additional assumption that the density / is continuous. Conjecture: Let X\,..., Xn,... be i.i.d. from a distribution F with a continuous density / , and let Z\,..., Z„,... be i.i.d. from an exponential distribution with mean 1, independent of the X;'s. Then the empirical spacings distribution S„(t) converges to the same limiting distribution as the empirical distribution from the i.i.d. sample Z\jf{X{),... ,Zn/f(Xn).

44

Heuristic Argument: By the mean value theorem F(X(i+1)) for some X^

= F(X(i))

< Vi < X(i+iy

+ f(Vi)(X(i+1)

-

X(i))

Thus

F ( X ( j + 1 ) ) - F ( X ( i ) ) = /(V5)r i .

(2)

However, since F{Xi) has a uniform distribution, the left side of (2) gives the spacings of an i.i.d. sample from the uniform distribution. Thus, the joint distribution of n/(y1)y1,...,n/(y„)yn will approach that of Z\,..., Zn. For large samples, the order statistics will approximate the appropriate quantiles, as will the Vi's sandwiched between them. Thus, the joint distribution of nY\,... ,nYn will approximate that of Zi/f{Xw),..., Zn/f{X(n)). The random function

5;c) = ^iEJ(t.oo)(2r*//(x w )) > i=l

on the other hand, will have the same distribution as 1

71 — 1

Sn(t) =

£

J(tl0o) (Zi/fiXi))

,

(3)

»=i

which is the empirical survivor function Z1/f(X1),...,Zn/f(Xn).

from the i.i.d. sample

A rigorous proof of the conjecture will depend first on a careful specification of the nature of the putative asymptotic equivalence between (1) and (3). A strong result would obtain if it could be shown that the stochastic process TJU (§n(t) — S(t)) converges to the same limit as y/n (Sn(t) — S(t) J. The asymptotic spacings distribution oo

/

e-tfWf(x)dz

(4)

-oo

is just the distribution of the random variable Z/f(X), which, by conditioning on X, can be viewed as a mixture of exponential distributions.

45

Example 1. If F is exponential, then the spacings themselves are exponential with E{Yi) = l/(ra — i) and the distribution of Sn can be calculated explicitly for finite n. We have n-l

E (sn(i)) = (n - I)""1 Y,

Pr

{ ^ > *M

i=l n-l

(n-l)-1Y,e~in~i)t/n

=

i=l

_ 1 - e x p ( - ( l - l/n)t) (n-l)(e'/n-l) from which we can see

that E (§n(tj) -y (1 — e ')/£ as required by (4).

Example 2. Consider a random variable with density function k

/(*) =

0 < a; < fc + 1

:+ l

where k > 0 is a parameter. (These distributions are members of the Beta family, rescaled for ease of computation.) Making the change of variable v-t

fc+1

we can calculate rfc+i

S(t) = / Jo

exp - t

_ 1 + 1/fc ti+i/k

r(2 +

fc + 1

fc + 1


Te-V

~vv^k dx Jo i/k)G1+1/k(t)

tl+l/k

where G 1+1 /fc(t) is the Gamma cdf with shape parameter 1 + 1/fc. These two examples illustrate ASD's with tails behaving like negative powers. Such distributions have at most a finite number of moments. In

46

particular, the expected asymptotic spacing length will be

JO

r

/>oo

roo

/

S(t)dt=

/ JO

e-tfWf{x)dxdt

/ Jf(x)>0

= /

f(x)/f(x)dx

Jf(x)>0 — 00

unless the support of / is bounded. Thus the distribution of an unbounded random variable, such as the Normal, cannot have an ASD with a tail 0(t~a) with a > 1 for then its expectation would be finite. 3

Characterization of asymptotic spacings distributions

Note that the ASD is in fact a Laplace transform. If X has density / , consider the random variable W = f(X). Then S{t) = E{e~tw). If / is monotone, then the density of W is given by /°/_1(w) | / ' o / - » |

w =

| / ' o / - » |

provided the density is differentiable. (If / is not monotone, then the density of W must include contributions from the components of f~1(w)). Thus, for example, if X is Exponential, then W is uniform with Laplace transform (1 — e~')/i as seen previously. If X is Uniform, W is degenerate at 1, with Laplace transform e~l. The representation of the ASD as a Laplace transform allows one to use Tauberian theorems 4 which relate the tail behaviour of the Laplace transform with the behaviour of the distribution near the origin. Specifically, if L is the Laplace transform of a distribution F, then as F(t)~L(l/t)/T(p+l) where [mlog(F(tx)/F(t))

i->0

_

log X

We see that what controls the tail behaviour of the ASD is the behaviour near the origin of W, which is determined by the way that the density / approaches the axis. It can readily be seen that if f(x) behaves like 0(xk) as x -> 0, then W will have a density like 0{w1/k) as w - • 0. The cdf of W will then behave like 0{wl>k+l) and the ASD will have a tail like 0 ( H 1 + 1 / f c ) ) .

47

Quantiles of asymptotic spacing distributions < O

"

CM


"

"^^ r i ^ 1 1 **«-^

~ ——

< o

^--^^^^x

3? CD Q. Q. Z>

r

o

—

Normal Exponential

-—

Beta(1,2 ' Uniform

1

CM

< O

CO


0.005 0.010

0.050 0.100 Probability

0.500 1.000

Figure 1. Upper quantiles of asymptotic spacings distributions from selected parent distributions

Obviously, the distribution of W will be unaffected by location shifts or reflections, and will be reciprocally scaled by rescaling of X. Thus the tail behaviour of the ASD is the same for all distributions in a location-scale family. If X is Normal, then the density of W is messy, but after some calculation can be seen to be approximately proportional to (— logw) - 1 / 2 near the origin. This does not behave like a power of w, indicating that the ASD will not have a negative-power tail. By making the change of variable s = — logw, we can see that the cdf of W will be approximately proportional to -!/2 f

/ .— log w

'da.

It follows that the tail of the ASD will be like /•OO

/

'

-l/2„-s

ds,

J log*

i.e., like the tail of exp(C) where C has a xl distribution.

48

0.01

0.10

1.00 10.00 Normal spacings quantiles

100.00

Figure 2. Q-Q-plots of Normal spacings

4

Numerical results and simulations

The quantiles for the ASD corresponding to given parent distributions can be computed numerically. These are plotted in Figure 1 for distributions with different tail behaviours. One can examine the suitability of the ASD for describing the empirical distribution of spacings for large but finite sample sizes. Figures 2 and 3 present Q-Q-plots for spacings of samples 1000 i.i.d. observations from the Normal and Exponential distributions, respectively, plotted against the respective ASD. These plots demonstrate a reasonably good fit. The plot is presented on a log-log scale to accommodate the heavy-tailed distribution. Note that on such a scale, a well-fitting plot should lie on a straight line with unit slope. Such a line is superimposed on the points, constrained to pass through the empirical upper 10 percentile.

49

0.01

0.10 1.00 10.00 Exponential spacings quantile

100.00

1000.00

Figure 3. Q-Q-plots of Exponential spacings

Acknowledgments I want to thank Leon Katz for raising the question that spawned this paper. I am indebted to Richard Lockhart and Michael Stephens for useful discussions on this topic. This research was supported by a grant from the Natural Sciences and Engineering Research Council (NSERC) of Canada. References 1. 2. 3. 4.

L. Katz, Personal communication, 2000. R. Pyke, J. Roy. Stat. Soc. 27B, 395 (1965). J. R. Blum and L. Weiss, Ann. Math. Stat. 28, 242 (1957). W. Feller, An Introduction to Probability Theory and its Applications v. 2, 2nd ed. (Wiley, New York, 1971).

T H E O R E T I C A L A N D COMPUTATIONAL ISSUES I N B A Y E S I A N ANALYSIS OF MULTIVARIATE O R D I N A L CATEGORICAL DATA W I T H R E F E R E N C E TO A N OPHTHALMOLOGIC S T U D Y A T A N U BISWAS Applied

Statistics

Unit, Indian Kolkata E-mail:

Statistical Institute, 700 108, India [email protected]

203 B. T.

Road,

Bivariate or multivariate ordinal categorical data is quite common in different real life situations. Several frequentist's approaches are available for the analysis of such data. In almost all those approaches, the computational burden is tremendous. As an alternative, some Bayesian latent variable based approach were suggested by some authors. In such cases also, the computation is a key issue. The computational burden of such Bayesian analyzes can be remarkably reduced by using the software WinBUGS. The present paper discusses one real life dataset on an ophthalmologic study, called the Wisconsin Epidemiologic Study of Diabetic Retinopathy (WESDR), which provides bivariate ordinal categorical d a t a with several eye-specific and subject-specific covariates. The techniques of the available Bayesian analyzes with reference to the WESDR data are discussed in the present paper. Some results on Bayesian model selection for this data is also discussed. By the help of some exploratory d a t a analysis an appropriate model is selected, and it is then analyzed in the Bayesian semiparametric way taking a Dirichlet process prior for the random effect (in the frequentist's sense). The final result on the analysis of WESDR data is presented. Several computational issues in this context are also discussed. Finally, the applicability of the present approach to some more general situation are also pointed.

1

Introduction

In several social, psychological and biomedical studies the response variable is ordinal categorical in nature, sometimes they are ordinal due to the absence of well-defined non-invasive direct measurements {e.g., mild, moderate, severe, etc.). If the response is of multivariate nature and each component of the response is ordinal categorical, we have multivariate ordinal categorical data to deal with. This kind of data are available in many real life situations. Ever since Dale 7 proposed the analysis of bivariate ordinal categorical data, a lot of subsequent studies were carried out in this interesting and important research area. Much of the early works were done on the frequentist's view point. Molenberghs and Lesaffre 27 used a multivariate Placket distribution as an extension of Dale's 7 model. These likelihood methods are, of course, computationally extensive. As an alternative, Williamson et al. 37 considered

50

51

the generalized estimating equations (GEE) approach. They fitted cumulative probit margins and a global odds ratio association model. Williamson et al. 38 discussed the applicability and usefulness of their computer programs GEECAT (generalized estimating equations for categorical data) and GEEGOR (generalized estimating equations using global odds ratio). Kim 19 carried out a latent variable based estimation technique for a bivariate ordinal data of an ophthalmologic study called the Wisconsin Epidemiologic Study of Diabetic Retinopathy (WESDR). Kim 19 used the Newton-Raphson iteration technique to find the estimates of the underlying parameters including the typically unknown cut-off points. But he admitted the computational difficulty of the technique in a more general set up. Williamson and Kim 3 6 , also with reference to the WESDR dataset, considered bivariate latent variable regression model using global odds ratio, which required no specific choice of underlying latent distribution except its continuity, and no specific structure of the correlation. A quasi-Newton method in a full maximum likelihood procedure was employed for the estimation of the model parameters. Kim et al. 20 discussed regression models for bivariate ordered categorical data from ophthalmologic studies. Some statisticians wanted to look at the multivariate ordinal categorical data from a Bayesian philosophy and tried to apply the Bayesian theory for analyzing such a multivariate ordinal categorical data with covariates. It is observed that, besides the difference in philosophical perspective, the Bayesian computation becomes more tractable and one may arrive at good solutions after some easy computational effort. In the present paper we discuss some of the available approaches and discuss some possible new directions, mainly with reference to a real life bivariate ordinal dataset obtained from the WESDR study. We discuss the WESDR study and the nature of the resulting bivariate ordinal data briefly in the next section. In Section 3, two available analyzes on WESDR data are briefly indicated. A discussion on appropriate model selection is provided in Section 4. In Section 5, a Bayesian analysis of the WESDR data using the selected model from Section 4 is carried out. The results are briefly discussed. Some important computational issues are discussed in Section 6. Finally, Section 7 ends with some concluding remarks.

52

2 2.1

WESDR Description of the WESDR

It was a population-based study conducted in Southern Wisconsin between 1980 and 1982 by Dr. Ronald Klein and his medical colleagues at the University of Wisconsin and supported by the National Eye Institute, NIH. In the baseline study, a total of 996 insulin-taking, younger onset diabetic patients were examined using standard protocols to determine the prevalence and severity of diabetic retinopathy and associated risk variables. The population of the study consisted of a probability sample selected from 10135 diabetic persons who received primary care in an 11-county area in southern Wisconsin from 1979 to 1980. A detailed description of the population can be available in Klein et al. 2 1 . Of the younger-onset persons (less than 30 years of age), 996 participated in the baseline examination (1980 to 1982). Subsequently there were two follow-up examinations on the same population after 4 and 10 years (cf. Klein et al. 24 ; Klein et al. 2 2 ) , with some missing data in this course of study. The baseline and the two follow-up examinations were performed in a mobile examination van in or near the city where the participants lived. The ocular and physical examinations included taking stereoscopic color fundus photographs of seven standard fields. The basic goals of the study (cf. Klein et al. 25 ) were • To find the associated risk factors which are important in planning a well-coordinated approach to the public health problem posed by the complications of diabetes (cf. Hamman 17 ; Rand 3 0 ) . • Identifying the possible patients at high risk level of severe retinopathy was quite important for advising ophthalmologic care. • Planning future studies such as controlled clinical trials of treatment of diabetes and diabetic retinopathy (cf. Rand 30 ; Palmberg et al. 2 8 ) . • Progression of diabetic retinopathy over time and associated factors was also an important aspect of study (cf. Wahba et al. 3 3 ). 2.2

Data description

WESDR data is a bivariate ordinal data. The retinopathy scale (RS) provided in the dataset is a more current one than the one used in some earlier works. Both the right and left eye retinopathy severity levels are recorded as two components of the bivariate response. The possible values of the retinopathy severity levels are 10, 21, 31, 37, 43, 47, 53, 60, 61, 65, 71, 75 and 85, corresponding to increasing levels of severity of retinopathy within an eye. A commonly used grouping is 10, 21-37, 43-53 and 60-85, which correspond to

53

no retinopathy (category 0), mild nonproliferative retinopathy (category 1), moderate to severe nonproliferative retinopathy (category 2), and proliferative retinopathy (category 3), respectively. Such a grouping is usually done for easy interpretation of the data, and it reduces the computational burden to a great extent. Most of the available works are done using this grouping. Three eye-specific covariates are recorded separately for each eye. The first one is the presence or absence of macular edema (ME), which is the effusion of serious fluid into the intersects of cells in tissues. The right and left eye refractive error (RE) in diopters, which is the ability of the eye to refract light which enters into it so as to form an image on the retina, is also recorded. The values can be negative or positive, negative values represent myopia (nearsightedness), and positive values represent hyperopia (farsightedness). In addition, the right and left eye intraocular pressure (IOP) in mmHg. is also measured. In addition there are 11 person-specific covariates. The first two are age at diagnosis (AgD) of diabetes in years and duration of diabetes (DuD) in years. To get current age at the time of examination, one has to add AgD and DuD. The other person-specific covariates are glycosylated hemoglobin (GH) in percent (a measure of control of blood sugar where lower values are considered better), systolic and diastolic blood pressures (SBP & DBP) in mmHg., body mass index (BMI) in kilograms per meter squared (using weight and height), pulse rate (PR) in beats per 30 seconds, sex, urine protein (UP) (present/absent), doses of insulin (DI) per day and area of residence (AR) (urban/rural). 3 3.1

Bayesian Analyses Bayesian Analysis of Baseline Data

The first major Bayesian analysis of the WESDR (baseline) data is due to Biswas and Das 2 . They considered 691 observations with full response and covariate information for their analysis. Suppose yn and ym be the responses from the left and right eye, respectively, for the it\\ individual. Both y^i and ym can take values 0, 1, 2 and 3. It is assumed that there are some underlying latent variables which are unobservable, but in effect we observe these y^ and ym- Underlying latent variables y*Li and y*Ri are postulated, which are responsible for observing y^ and ym in the following way: Vhi = 3 if y*Li S (7ij,7ij+i], 3 = 0,1,2,3, VRI =HfyRi G (72i,72/+i], Z = 0,1,2,3,

54

where 71/s and 72*'s are typically unknown cut-off points such that 7so < 7*1 < • • • < 7s4 f° r 5 = 1,2. Some mild conditions on these 71/s and 72;'s, such as (1) 7 i o = 720 = - 0 0 ,

(2) 711 = 721 =

a

714 = 724 = 00,

known constant,

are needed for identifiability (see Chen and Shao 3 , for details). The latent vector y* = {y*Li,y*m)T is assumed to follow a bivariate normal distribution N2{XiP, S). The joint distribution of yt = (yLi, VRi)T, yVs a r e written as 7r(i/?, j/i, i = !,•••, ^1)8, E, 7) N

n

5 3 lioi x J ( » £ i

e

(7ij.7ij+i],I/m G (721,721+1])

3,l€S

xN2(Xip,i:),

(1)

where N = 691, the number of individuals under study; 7 is the collection of all unknown 7^'s and 7 2i 's; S = {{j, I) : j = 0,1, 2, 3; I = 0,1, 2, 3}; I(X e A) = 1 or 0 according as X € A or not; and 1J, = 1 or 0 according as y* = (j, l)T or not. Without sufficient prior knowledge, noninformative priors for /?, S and 7 are taken. The conditional posteriors of the parameters are of known form. The computations can be carried out using Markov Chain Monte Carlo (MCMC) technique. In particular, posterior summary statistics of all the parameters are obtained using the software WinBUGS. The analysis shows that the severity of retinopathy among the younger onset diabetic persons is directly affected by DuD and DBP. Also GH was one vital covariate for retinopathy. 3.2

Analysis of the Baseline and 4-year Follow-up Data

Das and Biswas 9 considered the analysis of bivariate ordinal data repeated over time. They took the WESDR data for two time points only (baseline and 4-year follow-up data). Some notational adjustments are to be needed from the subsection 3.1. In place of yLi, yRi, y*Li, y*Ri, 1},, Xt we just write yLit, ymt, yliti y*mv !}*> -Kit to indicate that they are for time point t, t = 0,1. For the baseline data we write t = 0 and for the 4-year follow-up data we write are t = 1. The joint distribution of yit = {yLit,VRit)T, y*t = (ylit,y*Rit)T,s now written as T(j/*t. Va,i = 1, • • •, n, t = 0, l|/3, S , 7 )

55 n

1

= IIII

£

J

i*

x /

(»£«

G

(7ii. 7 y + i ] , J/ffit G (72J, 721+iD

i = i t = o |_j,ies

xA^X^ + Z^E),

(2)

where the sample size n is now 548, as some of the individuals of the baseline study were missing in the follow-up study. A random effect model (in the frequentist's sense) is assumed for y*t as y*t = Xit(3 + Zitbi + eit, where &;, a ^-dimensional vector, is the ith cluster (subject) effect, which is responsible for the longitudinal dependence. Here e;t is assumed to be bivariate normal with mean vector 0 and dispersion matrix E. For likelihood identifiability, E needs to be a correlation matrix (see Chib and Greenberg 5 ; Chen and Shao 3 ) . The prior for (3 is taken as Np(Po, Eo). Without sufficient prior information a relatively vague prior is chosen by setting /30 = 0 and Eo = 10 4 x identity matrix. This ensures the posterior to be driven by data. A noninformative prior for 7 is taken. A prior proportional to the normal density (with known mean po and known variance T _ 1 ) in the domain ( — 1,1) is taken for p, the only correlation parameter in the 2 x 2 correlation matrix E. A Dirichlet Process (DP) prior for the unknown distribution G of 6,'s are taken (see Ferguson 12 ; Kleinman and Ibrahim 2 6 ) . The software WinBUGS is used for the Markov Chain Monte Carlo (MCMC) computations. Metropolis-Hastings algorithm (Hastings 18 ) was employed for sampling from the posterior of p, as it becomes non-standard; for other parameters the popular Gibbs sampler (see Geman and Geman 16 ; Gelfand et al. 13 ) is used. Note that in case E is a proper dispersion matrix, a suitable inverted Wishart prior could be taken for E. In that case the Bayesian MCMC computations could be same, possibly requiring some smaller time as the conditional posterior of E would be inverted Wishart also. 3.3

Present Analysis Using all the Data

The WESDR data provides information of only baseline values (and not of the 4-year follow-up) on N — n = 143 individuals. Das and Biswas 9 ignored this information. Now this data on 143 individuals can be successfully used to get more information. The likelihood thus becomes n(y*t,yit, i = 1, • • •, n, t = 0,1; y*0,yio, i = n +!,•••, N\0, E, 7)

56 n

1

nn

J2

*n

S

l

%

^

x

^Lit

X J

^

G (7y.7y+i],»fl« G (72(,72i+i])

i 0

G

(Ty'7ii+i],yfliO

G

(72i,72i+il)

i=n+l

xAr2(XiO/3 + Z i 0 &i,S)},

(3)

after modifying (2). Our ultimate goal is to analyze the baseline and 4-year follow-up data by making full use of the data after suitable prior elicitation and model selection in terms of covariate selection. These issues are discussed in the next section. 4

Prior Elicitation and Model Selection

Most of the priors considered so far are vague priors. An empirical Bayes procedure could be carried out if some past data were available. For example, Angers and Biswas l considered the analysis of the 4-year follow-up data only. They used the baseline data to construct priors for different parameters. Some sensitivity analyses were also carried out by Angers and Biswas 1. But, in the absence of such past information, the vague priors are appropriate. In such a case, the posterior will be driven by the data. In fact, we have a quite large dataset, and the data should dominate the posterior in any case. In the context of the 4-year data of the WESDR, model selection in terms of covariate inclusion was carried out by Angers and Biswas l . They computed the standardized Bayes estimators and ordered them. Then the model with highest marginal probability is selected. It was observed in their study that the baseline values, DBP and GH are important covariate in different situations. They included these covariates in their model to analyze the data. But if we carry out our analysis simultaneously for the baseline and 4-year follow-up data, there is no question of taking the baseline values as covariates. The covariates which are not important can be excluded from considerations to ease the computational burden. Wahba et al. 3 3 carried out their analysis on a subgroup of the younger onset population, consisting of 669 subjects with no or nonproliferative retinopathy. They considered retinopathy scale (RS) as the response variable. Thus they reduced the bivariate ordinal data problem to a simpler univariate ordinal data problem. Some exploratory GLIM modeling using the SAS procedure LOGISTIC (SAS Institute 31 ) were carried out and after some exploratory

57

considerations they took the model. Wahba et al. 3 3 observed that the effect of GH was very strong and fairly linear in the logit and that the effects of AgD, DuD and BMI were strong and nonlinear. We have done a similar detailed exploratory study using SAS with the baseline and 4-year data. We observe that the effect of DuD in a log-scale explains the data in a better way. Note that Wahba et al. 3 3 also observed that DuD has a nonlinear effect. We also find that the effects of GH and DBP are significant in the context of retinopathy levels. In fact, Klein et al. 23 reported that GH is a strong predictor of progression of diabetic retinopathy in the younger onset group. Angers and Biswas * also observed the same scenario. DBP came out as an important covariate in the study of Kim 19 and Biswas and Das 2 . In the present study we observe possible interaction between GH and DBP. Thus the model has to be taken with care. This is discussed in Section 5. 5

Present Bayesian Analysis W i t h the Selected Model

We now carry out a Bayesian semiparametric analysis of the baseline and 4year follow-up data of the WESDR. We take the selected covariates (discussed in Section 4) into the model, suitable transformations and interaction effects are also considered following our exploratory study described in the previous section. This seems to be a correct model which can explain the retinopathy levels in terms of different covariates. We use all the available data, i.e., the likelihood is given by (3). We impose an additional restriction 71 j = ^2j for all j , which seems logical as the two eyes should behave in the same way. The random effect model (in the frequentist's sense) is justified from intuitive feelings. Besides the significant covariates taken in the model (after excluding the insignificant ones) there must be several unobserved (most of which are unobservable) factors that can be responsible for retinopathy and also the association between the retinopathy levels between two eyes. As the two eyes of the same human being are considered, several nerves, tissues, same reading habit and work habit, same environmental exposure, etc. are responsible for similar effects on both the eyes. These unobservable factors are expressed in terms of the uncertainty 6j's. The model is then taken as y*kit = constant

+ \og(DuD) x j3DuD + GH x (3GH + DBP x (3DBp

+ (GH x DBP)

x PGHXDBP

+ bki + ekit,

k =

L,R.

A Bayesian semiparametric study as in Das and Biswas 9 was carried out with different choices of the prior. The prior for f3 is taken as a vague prior. This is taken by setting (3 ~ Np(0, EQ), where So = 10 4 x identity matrix, i.e., by

58

setting the variances to be large. Some noninformative prior proportional to unity is set for 7. The prior for p is the same discussed in subsection 3.1. A Dirichlet process prior is taken for bi = (&Lz!&JRi)T's such that bi ~ G ,

G~DP{a,G0), with Go ~ iV m (0,r), with some known T. When G 0 is known, i.e., G has a prior probability model with known hyperparameters, the posterior distributions can be easily obtained as in Wilks et al. 35 . Here we have set F = identity matrix. The posteriors are observed to be dominated by the data. We primarily take priors of the form discussed in Section 3. Now we discuss the nature of the conditional posteriors without providing the exact mathematical expressions. The conditional posterior of /3 is multivariate normal. The conditional posterior of p is non-standard, to sample from it one has to use the Metropolis-Hastings algorithm (see Hastings 1 8 ) . With some probability a(p, p ) (easily computable) we move to a candidate value p from the present value p, and with probability 1 — a(p, p ) we stick to p. We take p = p + h, where h is a random zero mean increment. The variance of h is usually taken to be of order O(^). The conditional posterior of 71J is found to be uniform over some domain. The conditional posterior of y*t is truncated bivariate normal, truncated in some rectangular region. Finally, the conditional posterior of bi is a mixture where one piece is normal and the others are point masses. That is, with some probability we choose bi = bj and otherwise we sample from the normal density. For detailed mathematical forms of the conditional posteriors, one can see Das and Biswas 9 . The derivation of the present case is similar to Das and Biswas 9 with a slight possible change in some expressions. The posterior summary statistics of the relevant parameters are then obtained. From our final analysis, the posterior mean of PDUD is 0.78, that for POH is 0.52 and for PDBP is 0.32. The posterior mean of PGHXDBP is 0.27. Thus all of DuD, GH and DBP have positive effect to the retinopathy on both the eyes in the sense that presence of any of them increases the retinopathy levels. The interaction of GH and DBP is also affecting the retinopathy in a positive way. The posterior mean of p, the polychoric correlation (correlation between two categorical random variables) is observed to be 0.862. Thus, quite a high correlation between the two eyes is present. This justifies the need for such an analysis of multivariate categorical data - one would loose much information if the analysis were carried out marginally for each response, separately. It is also observed that the posterior standard deviations of the regression coefficients are low for most of the cases. The Monte Carlo

59 error (MC error) is also quite small, these are less than 0.001 for most of the parameters. In turn we have tried the same analysis using logistic distributions for cm and emt (see Qui et al. 29 , for such logistic error distributions). But the results are almost same as that for normal errors. Hence we are not discussing the figures separately. 6

Computational Issues

In such analyses, computations play a major role. The main objection to the earlier frequentist's solutions (without going to the philosophical issues) was their tremendous computational burden. In fact, in some more complicated scenarios, most of those approaches become computationally intractable. Thus, along with the philosophical view point, we look at the computational burden of the Bayesian approaches with great interest. Qui et al. 29 discussed some such computational issues for multivariate ordinal data in some other situations. Their case was much simpler in the sense that they did not consider complicated random effect distributions like the Dirichlet process semiparametric model. Moreover the prior for p in our case yielded a non-standard conditional posterior, requiring Metropolis-Hastings algorithm to be carried out. All the computations of Biswas and Das 2 and Das and Biswas 9 were carried out using MCMC technique which avoids the evaluation of high dimensional numerical integration. The computations in Angers and Biswas 1 was done using a computer program in S-Plus. It took hours for the computation for model selection. The MCMC requires intensive computation and careful assessment of convergence. For computations, the freely available software WinBUGS is used and it eased the computational burden greatly. WinBUGS is a window version of the software BUGS (Bayesian analysis Using Gibbs Sampler), developed by MRC, Biostatistics Group, Institute of Public Health. For details see http://www.mrc-bsu.cam.ac.uk/bugs/ or the manual of WinBUGS 1.3 (see also Spiegelhalter et al. 3 2 ) . To operate the necessary computations, the software requires the likelihood, prior and data, although we have mentioned the distribution of the conditional posteriors only to visualize what is going on behind the tandom. It generates random samples from a series of conditional posterior distributions specified in the Bayesian model according to an MCMC algorithm. The analysis of the baseline data in Biswas and Das 2 was initially programmed in C. But it took more than 24 hours for the computation. By WinBUGS the computations were done in less than 4 hours. For the DP

60 prior with baseline and 4-year follow-up data in Das and Biswas 9 , WinBUGS took about 16 hours for the computation. After selecting the model and exclusion of several insignificant covariates, for the present analysis WinBUGS took about 12 hours for each computation. Standard approaches can be considered for assessing convergence (Cowles and Carlin 6 ) and for model checking (Gelman et al. 14 ) and comparisons (Chib 4 , Diciccio et al. n ) . In the present paper the convergence of Gibbs sampler is ensured by the basic approach of Gelman and Rubin 15 . Starting from some initial values of the parameters we generate 4 chains of values, each chain being generated starting from an over-dispersed distribution and with a sample size of 8000. We delete 4000 replications as "burning" samples to minimize the effect of initial values and retain the values of the next 4000 replications to approximate the posterior distribution. To monitor the convergence we focus our attention to the parameters of interest, namely DuD, GH, DBP and GH x DBP. Following Gelman and Rubin 15 , we compute the between and within chain mean squares of the retained values, say B and W respectively, for each of the parameters. Then we find s2 = (4000 - l)W/4000 + 5/4000, i/ = s2 + (4x 4000) _1 S, and finally the potential scale reduction factor r = v/W. The potential scale reduction factors are nearly 1, and this suggests that the desired convergence is achieved in the Gibbs sampler. A 4000 updates were burn in as the initial samples, followed by a further 4000 samples which were used to obtain the posterior summary statistics like the posterior mean, median, standard deviation, MC error, 95% probability interval for each of the parameters under consideration. 7

Concluding Remarks

• It could be of interest to analyze the 10-year follow-up data at the same occasion. But we could not access the data. • In addition to the probit link, a Iogit link is also tried, of course with not much difference in the results. This result was somewhat expected, as there is not much difference in the tail behavior of the normal and the logistic distributions. With the help of WinBUGS the computation is easily doable, and it only needs a slight change in the program. • Some nonparametric technique in terms of smoothing spline ANOVA was employed by Wahba et al. 33 . We are now trying to employ a suitable

61

wavelet approach. This is under consideration and will be pursued in a future communication. • If we had to deal with multivariate data, the number of correlation components in the correlation matrix would be greater than one, namely Pi2,Pi3, • • •• In this case writing p = (pi2,Pi3, • ' ' ) T i t n e prior of the kdimensional vector p (say) can be taken to be proportional to a fc-variate normal density (with known mean vector and dispersion matrix) truncated in a set C which is a convex solid body, subset of [—1, l] fc that leads to a proper correlation matrix. See Chib and Greenberg 5 for details. One can set the variances large to make it an uniform prior. If, instead, S were taken as a proper dispersion matrix (unknown), an inverted Wishart prior could be appropriate. The conditional posterior would be inverted Wishart in that case. • One relevant point may be the case where some component (s) of the multivariate categorical response is (are) not ordinal. In that case one can possibly use the Rasch model for that component(s). The problem of combining it in our present set up is a challenge. Latent variable approach will not be applicable for those components. The theory is under study with some ordinal components and some (non-ordinal) only categorical components. The details will be pursued in a separate communication. Also one interesting situation could be the case where some of the components of yi, say yu, • • • ,ysi are ordinal categorical, and the remaining components of j / i ; say ys+i,i, • • • ,VH are continuous. See Das et al. 10 for a Bayesian approach in such a situation. • If, in the multivariate case, some of the cells of the fc-way ordered classification (two-way, for bivariate case) are empty, the analysis will become more complicated. In a bivariate case, such a situation is observed and analyzed by Weiss 34 in the frequentist's set up. But Weiss 34 observed that the log-likelihood function is not globally concave resulting serious difficulty in estimation. In the Bayesian set up, the likelihood will be simply as (1) with an indicator multiplied for each individual i, which will be 1 if the bivariate observation vector belongs to the non-empty cells, and will be 0 otherwise. A Bayesian analysis in this direction has been done by Das and Biswas 8 which shows that the Bayesian solution does not have the drawbacks of the frequentist's approach. • Finally, we would like to provide some motivation of the random effect (in the frequentist's sense) semiparametric model. We are, in fact, observing some covariates. But, there is reason to believe that there are numerous physiological factors such as several nerves are affecting the retinopathy of an individual, most of them are unobservable. To model their effects in the retinopathy level, one has to bring some random component (in the fre-

62

quentist's sense) under consideration. See Das and Biswas discussion in this context.

9

for a detailed

Acknowledgments The author wishes to thank the two referees for their careful reading and some suggestions which led to some improvement over an earlier version of the manuscript. The author wishes to thank Drs. Ronal Klein and Barbara E. K. Klein of the University of Wisconsin for providing both the baseline and 4-year follow-up data of the WESDR study. The WESDR project was originally supported in part by grant EY 03083 (R. Klein) from the National Eye Institute, NIH. References 1. J.-F. Angers and A. Biswas, Technical Report. (University of Montreal, CRM 2757, 2001). 2. A. Biswas and K. Das, Statistics in Medicine 2 1 , 549 (2002). 3. M.-H. Chen and Q.-M. Shao, Journal of Multivariate Analysis 7 1 , 277 (1999). 4. S. Chib, Journal of the American Statistical Association 90, 1313 (1995). 5. S. Chib and E. Greenberg, Biometrika 85, 347 (1998). 6. M.K. Cowles and B.P. Carlin, Journal of the American Statistical Association 89, 883 (1994). 7. J.R. Dale, Biometrics 42, 909 (1986). 8. K. Das and A. Biswas, Technical Report (Applied Statistics Division, Indian Statistical Institute, 2000). 9. K. Das and A. Biswas, Technical Report. (Applied Statistics Division, Indian Statistical Institute, 2001). 10. K. Das et al., Calcutta Statistical Association Bulletin 49, 255 (1999). 11. T.J. Diciccio et al., Journal of the American Statistical Association 92, 903 (1997). 12. T.S. Ferguson, Ann. Statist. 1, 209 (1973). 13. A.E. Gelfand et al., Journal of the American Statistical Association 85, 972 (1990). 14. A. Gelman et al., Statistica Sinica 6, 733 (1996). 15. A. Gelman and D.B. Rubin, Statistical Science 7, 457 (1992). 16. S. Geman and D. Geman, IEEE Trans, on Pattern Analysis and Machine Intelligence 6, 721 (1984).

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17. R.F. Hamman, Proceedings of the Diabetes Control Conference, Atlanta, Centre of Disease Control, 32, (1982). 18. W.K. Hastings, Biometrika 57, 97 (1970). 19. K. Kim, Statistics in Medicine 14, 1341 (1995). 20. K. Kim et al. in Collected Papers in Honor of Retirement of Professor Chung Han Yung. Seoul National University, Department of Computer Science and Statistics Alumni Association, eds. J.Y. Park et al, 36, (1996). 21. R. Klein et al., American Journal of Epidemiology 118, 228 (1983). 22. R. Klein et al, Arch. Ophthalmol. 112, 1217 (1994). 23. R. Klein et al., Journal of the American Medical Association 260, 2864 (1988). 24. R. Klein et al., Arch. Ophthalmol. 107, 237 (1989). 25. R. Klein et al, Am. J. Epidemiol. 119, 54 (1984). 26. K.P. Kleinman and J.G. Ibrahim, Biometrics 54, 921 (1988). 27. G. Molenberghs and E. Lesaffre, Journal of the American Statistical Association 89, 633 (1994). 28. P. Palmberg et al., Ophthalmology 88, 613 (1981). 29. Z. Qui, Journal of Biopharmaceutical Statistics 12, (2002) 30. L.I. Rand, Am. J. Med. 70, 595 (1981). 31. SAS Institute SAS/STAT User's Guide, Version 6, 4th ed. (SAS Institute, Inc., Cary, North Carolina, 1989). 32. D.J. Spiegelhalter et al., Technical report (University of Cambridge, MRC Biostatistics Unit, 1994). 33. G. Wahba et al, Ann. Statist. 23, 1865 (1995). 34. A.A. Weiss, Applied Statistics 42, 487 (1993). 35. W.R. Wilks et al, Biometrics 49, 441 (1993). 36. J. M. Williamson and K. Kim, Statistics in Medicine 15, 1507 (1996). 37. J.M. Williamson et al., Journal of the American Statistical Association 90, 1432 (1995). 38. J. M. Williamson et al., Computer Methods and Programs in Biomedicine 58, 25 (1999).

SECOND-ORDER MOMENTS A N D MUTUAL INFORMATION IN T H E ANALYSIS OF TIME SERIES DAVID R. BRILLINGER Statistics Department University of California Berkeley, CA, 94720-3860 E-mail: [email protected] A statistical network is a collection of nodes representing random variables and a set of edges that connect the nodes. A probabilistic model for such is called a statistical graphical model. These models, graphs and networks are particularly useful for examining statistical dependencies amongst quantities via conditioning. In this article the nodal random variables are time series. Basic to the study of statistical networks is some measure of the strength of (possibly directed) connections between the nodes. The use of the ordinary and partial coherences and of mutual information is considered as a study for inference concerning statistical graphical models. The focus of this article is simple networks. The article includes an example from hydrology.

1

Introduction

Science concerns relationships. The question that usually arises is what is the form of some relationship. A lesser question is how strong is a relationship. The work presented considers the use of partial coherency, and of coefficients of mutual information as measures of the strength of association of connections. An example involving river flows measured at a succession of dams along the Mississippi River is presented. Here the nodes are in series and the edges are directed. The locations of the dams are provided in Figure 1. Basic books discussing statistical graphical models include Cox and Wermuth 7 ,Whittaker 19 , Edwards 8 , Lauritzen 16 . The paper has the following sections: Mutual Information, Networks, Results, Discussion and Extensions. 2 2.1

Mutual Information Continuous Case

The field of information theory provides some concepts of broad use in statistics. One of these is mutual information. It is a generalization of the coefficient of determination, corr{X,Y}2, and it unifies a variety of problems. For a bivariate random variable (X, Y) with density function p(x, y) the

64

65

mutual information (MI) is defined as [ P(x,y) log P}X'y) dxdy Js px{x)PY{y) where S is the region p(x, y) > 0. As an example, for the bivariate normal the MI is given by IXY

=

IXY

=

(1)

-^og(l-p2XY)

where pxy is corr{X,Y}. The coefficient IXY has the properties of: 1). Invariance, IXY = luv if the transformation (X,Y) ->• (U,V) has the form U — f(X),V = g(Y) with / and g each differentiable 1-1 transforms. 2). Non negativity, IXY

> 0.

3). Measuring independence in the sense that IXY = 0 if and only if X and Y are statistically independent. 4). Providing a measure of the strength of dependence in the senses that i) IXY = oo if Y — g(X), and ii) Ixz < IXY if X is independent of Z given Y. The property 3) that IXY = 0 only if X and Y are independent stands in strong contrast to the much weaker correlation property of pXY. The estimation of entropy There are several methods that have been used. Nonparametric

estimate

Suppose one is considering the bivariate random variable (X, Y). Supposing further that p(x,y), is an estimate of the density p(x,y), for example a kernel estimate, then a direct estimate of the entropy is S2J2

P{iS,jS) log p{iS,j6) = E{\og p(X,Y)}

for 6 small.

In the same way E{log px(X)}, E{\og py(Y)} may be estimated and one can proceed to an estimate of the mutual information via expression (1). References to the type of entropy estimate just described and some statistical

66

properties include: Joe 15 , Hall and Morton 13 , Fernandes 9 , Hong and White, 14 , Granger et al. 12 . Difficulties with this form of estimate can arise when px(-), PY(-) are small. The nonparametric form also runs into difficulty when one moves to higher dimensions. A sieve type of estimate is presently being investigated for this situation, in particular an orthogonal function expansion employing shrunken coefficient estimates. Parametric estimates of entropy If the density p(x,y\8) depends on a parameter 9 that may be estimated reasonably then an immediate estimate of the entropy is provided by /

p(x,y\6)

log p(x,y\6)

dxdy.

Another form of estimate is based on the likelihood function. Suppose one has a model for the random variable (X, Y) including the parameter 0, (of dimension u). Suppose the model has the property that X and Y are independent when 6 = 0. When there are n independent observations the log-likelihood ratio for the hypothesis 6 = 0 is

Vlog-

p Xi Vi

( ' W

'Px{xi)p Y (yi)

with expected value nlxYThis suggests the use of the log-likelihood ratio statistic divided by n as an estimate of IXY- A further aspect of the use of this statistic is that its distribution will be approximately proportional to xl, where v is the dimension of 6, when X and Y are independent. Partial analysis When networks are being considered the conditional mutual information is also of use. For a trivariate random variable X, Y, Z one can consider 7IXY\Z

f f f = J

I

] J P(x,y,z)

\ 1

log

P(X1 Vl Z)P(Z)

, , ,

p{x^z)p{yjZ)

dxdydz

67

Its value for the trivariate normal is --log(l -

p2XY\z)

with PXY\Z the partial correlation of X and Y having removed the linear effects of Z. 2.2

Processes

A disadvantage of MI as introduced above is that it is simply a scalar. As consideration turns to the process case, i.e. functions, it seems pertinent to seeks to decompose its value somehow. 1. Time-side approach The entropy of a process is denned by a suitable passage to the limit for example as lim

E{logp(X1,X2,...,XT)}

T—foo

where p(xi, ...,XT) denotes the density of order T. To begin one can simply consider the mutual information of the values Y(t + u),Y(t) or of the values Y (t + u), X(t). This leads to a consideration of the coefficients IYY(U)

and

IYX(U)

i.e. mutual information as a function of lag u. References to this idea include: Li 17 and Granger and Lin n . 2. Frequency-side approach Similarly it seems worth considering the mutual information at frequency A of two components of a bivariate stationary time series. This could be defined as the mutual information of the spectral increments dZx(X) and dZy(A) of the Cramer representation. Because these variates are complex-valued a 4-variate random variable is involved. In the Gaussian case the MI at frequency A is - log(l -

\RXYW\2)

where \RXY (A)|2 is the coherence of X and Y at frequency A. The overall information rate is

- T i o g a - \R(U) \2)du, J — 7T

(see Granger and Hatanaka 1 0 ) .

68 Wisconsin

St Anthony Falls^ Dcrn No

River

Minnesota Iowa O

Lock & Dan

£

Reservoir

ft

Strean Gage

Figure 1. The locations of the 10 dams along the Mississippi River some of whose flow rates are studied.

69 In the general case for each frequency one might construct an estimate, /xy(A), based on kernel estimates of the densities taking empirical FT-values near A as the data. One way to estimate the MI, suggested above, is to fit a parametric model and then to use the log-likelihood ratio test statistic for a test of independence. A novel way, also being pursued, is to use first recurrence time estimates of entropy (see Ornstein and Weiss 18 and Wyner 2 0 ) . 3

Networks

In crude terms a network is a box (or node) and line (or edge) diagram and some of the lines may be directed. In our work a box corresponds to a random entity, to a random variable, to a time series or to a point process. In studying such models the methods of statistical graphical models provide pertinent methodology. Typically these models are based on conditional distributions. See the books by Edwards 8 , Whittaker 19 , Lauritzen 16 , Cox and Wermuth 7

If A, B, C represent nodes a question may be as simple as: Is the structure A -»• B -> C appropriate or is it better described as (A, B) -> CI On the other hand the question may be as complicated as: What is the wiring diagram of the brain? Figure 1 shows the locations of dams of a network along the Mississippi River. Since the bulk of the water flows south an elementary graphical model for this situation is: Dam 1 -> Dam 2 -> ... -> Dam 10. Of course there are other sources of water, such as entering rivers and rainfall to be taken note of. The figure and the data to be analyzed are taken from www.mvp-wc.usace.army.mil/projects/lockjdam.html One reference to the approach of this paper is Brillinger 3 . 4 4.1

Results Mississippi River Flow

The waters of the Mississippi River flow from Minnesota in the north of the United States to the Gulf of Mexico. Flooding along the river has long been a concern, and the U.S. Army Corps of Engineers has constructed a series of locks for flood control and as an aid to navigation. The waters flowing may be viewed as a system added to by precipitation and by flow from entering streams and runoff and reduced by evaporation, absorption and diversion. The basic data employed in the study to be described are the daily water

70

Coherence dams 2 and 4 1.0 0.8 0.6 -

Coherence dams 2 and 5 1.0

\ 0.8

\

0.6

\

0.4 •

V

0.4 0.2

0.2 0.0 -

0.0

WiiikiiiiiJjli ifmmNmtm 0.2

0.4

0.0

0.0

0.2

0.4

frequency (cycles/day)

frequency (cycles/day)

Coherence dams 4 and 5

Partial coherence, 25I4 1.0 • 0.8 • 0.6 0.4 • 0.2 • | L | „ , | L. . l l j . ^ i l l . . , . .

0.0 •

0.0

0.2

0.4

frequency (cycles/day)

FV'NMVFWL

0.0

.irawi 0.2

0.4

frequency (cycles/day)

Figure 2. Estimated partial coherence of log(flow rate) at Dams 2 and 5 given those at Dam 4. The horizontal line gives the approximate upper 95% null level.

71

Partial coherence, 35|4

0.0

0.2

0.4

frequency (cycles/day)

Partial coherence, 24(3

0.0

0.2

0.4

frequency (cycles/day)

Partial coherence, 25|34

0.0

0.2

0.4

frequency (cycles/day)

Figure 3. The estimated partial coherence of log(flow rate) at Dams 3 and 5 given Dam 4, Dams 2 and 4 given Dam 3 and Dams 2 and 5 given those at Dams 3 and 4. The horizontal lines give the approximate upper 95% null level.

72

flow rates as recorded at a succession of dams along the river. The data are daily from 1960 on and were obtained from the WWW site mentioned above. Figure 1, taken from that web site, shows the locations of the dams. Entering streams may be seen. Consider for example the logarithms of the flow rates, Y2(t),Yi(t),Y5(t) at Dams 2, 4, 5. Their locations may be seen in Figure 1. One sees the bulk of the waters passing from Dam 2 to Dam 4 and then onto Dam 5. One sees the Zumbo River entering between Dams 4 and 5 and the three other rivers entering between Dam 2 and Dam 4. The logarithms of the flow rates are taken as the basic variables. This situation will be studied as providing a useful test bed for studying the effectiveness of the partial coherence and mutual information parameters. Figure 2 presents the results of a partial coherence analysis focusing on Dams 2, 4, 5. The top right panel provides the estimated coherence functions of Dams 2 and 5. The horizontal line gives the approximate upper 95% null level under the null hypothesis of zero coherence. One sees high coherence at the low frequencies. The bottom right plot is the estimated partial coherence function of Dams 2 and 5 having removed the linear time invariant effects of Dam 4. Once again the horizontal line gives the approximate upper 95% null level under the null hypothesis of zero coherence. One looks for more than about 5% of the values lying above these lines. In the partial coherence case one sees not too much activity. In this situation the partial coherence could have been anticipated to be negligible because of Dam 4's being so close to Dam 5, i.e. highly effective in blocking off the direct effects of Dam 2. As an aid to understanding this analysis one can consider the model Y5(t)

=

I a{t - u)Yi(u)du

+ e{t)

Y4(t)

=

/ b(t - u)Y2(u)du

+ r)(t)

with e and 77 noise processes. The partial coherence estimated is then the coherence of the processes e and 77. The analysis is now extended to 4 series. Figure 3 provides the estimated partial coherences of Dams 3 and 5 having removed the effects of Dam 4 and of Dams 2 and 4 having removed the effects of Dam 3 and also that of Dams 2 and 5 having removed the effects of Dams 3 and 4. The required formulas may be found in Brillinger 2 . The networks contemplated in the analysis are the parallel and series ones. Logically the series structure is appropriate. All three of the partial coherences appear weak. This is consistent with the series structure of the graph as was anticipated. Had there been parallel links

73

Mutual information - Real parts

0.0

0.1

0.2

0.3

0.4

0.5

frequency (cycles/day)

Mutual information - Imaginary parts

0.6

0.4

0.2

0.0 0.0

0.1

0.2

0.3

0.4

0.5

frequency (cycles/day)

Figure 4. The top panel provides an estimate of the MI of the two real parts of the Mississippi river flows at Dams 2 and 5. The bottom panel similarly refers to the two imaginary parts. The dashed line gives the approximate upper 95% null level.

74

through Dams 3 and 4 instead of the serial ones then the partial coherences 35|4 and 24|3 would not be expected to be near 0 generally. As a further example the MI of series 2 and 5 was estimated as a function of frequency A, actually the Mi's of the two real parts of the frequency components and of the two imaginary parts. These estimates are graphed in Figure 4. Approximate 2.326 s.e. limits are obtained by randomly altering the phases of the empirical FT components, following the procedure of Braun and Kulperger 1. In the plots one notes some apparent association at the lower frequencies. This could arise, for example, from the occurrence of snow or rain storms affecting the segments of the river at the same time. An estimate of the full MI is currently being developed following the discussion of Section 2 . A point process analysis of this situation is developed in Brillinger 4 . These data are also considered in Brillinger 5 .

5

Discussion and Extensions

The estimated partial coherences of the dams' flow rates had the forms anticipated given the physical knowledge of the situation. The coefficient of mutual information is a unifying concept extending second-order quantities that have restricted applicability. Its being 0 actually implies that the quantities involved are statistically independent. Another important advantage is that the MI pays no real attention to the values of the process. They can be non-negative, integers or proportions for example. The MI is useful when one wishes to make inferences stronger than: "The hypothesis of independence is rejected" and more of the character "The strength of connection is I." During the work the plot of the function IXYW, appear more useful than simple scalars IXY- Both parametric model-based estimates and nonparametric estimates of mutual information have been mentioned and computed. A number of extensions are available and some work is in progress. One can consider the cases of: point processes, spatial-temporal data, local estimates, of learning, of change, of trend, and of comparative experiments. Brillinger 6 contains related ideas and examples from neurophysiology. One needs to develop the statistical properties of other estimates of MI such as the estimate based on the waiting time and the sieve estimates.

75

Acknowledgments Dr. Partha Mitra made some stimulating remarks concerning the use of mutual information. The Referees' remarks led to clarifications and reductions. I thank them. Part of the material was presented as the Opening Lecture at the XXXIeme Journees de Statistique in Grenoble in 1999. This research was supported by the NSF grants DMS-9704739 and DMS9971309. References 1. J. Braun and R. Kulperger, Commun. Statist.-Theory Meth. 26, 1329 (1997). 2. D. R. Brillinger, (1975). Time Series: Data Analysis and Theory (HoltRinehart, New York. Republished as a SIAM Classic in Applied Mathematics 2001). 3. D. R. Brillinger, Brazilian Review Econometrics 16, 1 (1996). 4. D. R. Brillinger, Technometrics 43, 266 (2001). 5. D. R. Brillinger, Proc. Interface 2001. 6. D. R. Brillinger, Revista Investigacion Operacional (to appear, 2002). 7. D. R. Cox and N. Wermuth, Multivariate Dependencies: Models, Analysis, and Interpretation (Chapman & Hall, London, 1998). 8. D. Edwards, Introduction to Graphical Modelling (Springer, New York, 1995). 9. M. Fernandes, Nonparametric entropy-based tests of independence between stochastic processes (Preprint, Fundacao Getulio Vargas, Rio de Janeiro, 2000). 10. C. W. J. Granger and M. Hatanaka, Spectral Analysis of Economic Time Series (Princeton University Press, Princeton, 1964). 11. C. W. J. Granger and J-L. Lin, J. Time Series Anal. 15, 371 (1994). 12. C. W. J. Granger, E. Maasouni and J. Racine, A dependence metric for nonlinear time series (Preprint, Dept. of Economics, UCSD, 2000). 13. P. Hall and S. C. Morton, Ann. Inst. Statist. Math. 45, 69 (1993). 14. Y. Hong and H. White, Asymptotic distribution theory for nonparametric entropy measures of serial dependence (Preprint, Economics Department, UCSD, 2000). 15. H. Joe, Ann. Inst. Statist. Math. 4 1 , 683 (1989). 16. S. L. Lauritzen, Graphical Models (Oxford University Press, Oxford, 1996).

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17. W. Li, J. Statistical Physics 60, 823 (1990). 18. D. S. Ornstein and B. Weiss, IEEE Inf. Theory 39, 78 (1993). 19. J. Whittaker, Graphical Models in Applied Multivariate Statistics (Wiley, New York, 1990). 20. A. J. Wyner, Ann. App. Prob. 9, 780 (1999).

SPATIAL ASSOCIATION B E T W E E N C O M M U N I T Y AIR POLLUTION A N D HEART DISEASE: ANALYSIS OF CORRELATED DATA

SABIT CAKMAK, RICK B U R N E T T Health

Canada, PL 3104E, Ottawa, On., K1A 1B6, E-mail: [email protected]

Canada

MICHAEL J E R R E T T School of Geography

and Geology, McMaster

University,

Hamilton,

Canada

M A R K S. G O L D B E R G Departments

of Epidemiology,

Biostatistics, and Occupational University, Canada

Health,

McGill

A R D E N P O P E III Economics

Department,

Brigham

Young

University,

Provo,

Utah,

USA

R E N J U N MA Department

of Mathematics

and Statistics,

University

of New Brunswick,

Canada

DANIEL KREWSKI Institute

of Population

Health,

University

of Ottawa,

Canada

Cohort study designs are often used to assess the association between community based ambient air pollution concentrations and health outcomes, such as mortality, development and prevalence of disease, and pulmonary function. Typically, a large number of subjects are enrolled in the study in each of a small number of communities. Fixed site monitors are used to determine long-term exposure to ambient pollution. T h e association between community average pollution levels and health is determined after controlling for risk factors of the health outcome measured at the individual level (i.e., smoking). We present a new spatial regression model linking spatial variation in ambient air pollution to health. Health outcomes can be measured as continuous variables (pulmonary function), binary (prevalence of disease), or time to event data (survival or development of disease). The model incorporates risk factors measured at the individual level, such as smoking, and at the community level, such as air pollution. We demonstrate that the spatial autocorrelation in community health outcomes, an indication of not fully characterizing potentially confounding risk factors to the air pollution-health association, can be accounted for through the inclusion of location in the deterministic component of the model assessing the effects of air pollution on health. We present a statistical approach t h a t can be implemented for very large cohort studies. Our methods are illustrated with an analysis of the American Cancer Society cohort to determine whether the prevalence of heart disease is associated with concentrations of sulfate particles.

77

78

1

Introduction

Cohort study designs are often used to assess the association between community based ambient air pollution concentrations and health outcomes, such as mortality, development and prevalence of disease, and pulmonary function. Typically, a large number of subjects are enrolled in the study in each of a small number of communities. Practical considerations of implementation motivate this epidemiologic design, in that it is relatively easy to recruit additional subjects in a community compared to identifying communities with appropriate longitudinal pollution monitoring records. Fixed site monitors are used to determine long-term exposure to ambient pollution. The association between community average pollution levels and health is determined after controlling for risk factors of the health outcome measured at the individual level (e.g., smoking). Standard statistical computing software programs (e.g., SAS 10 ) can be used for analysis if the assumption of statistical independence between subjects is appropriate. Health responses, however, often cluster by community, indicating that responses of subjects within the same community are more similar than responses of subjects in different communities. This implies that community itself poses some risk to health. Community-level variables, such as measures of socio-economic status of the community, can be used to model this unexplained risk in addition to individual-level risk factors. Failure to account for all the variation between community health outcomes even after controlling for individual and community level risk factors can lead to downward biased estimates of the uncertainty in the community-level risk factors, including air pollution (see Ware and Strom 1 3 ). Additional bias in the uncertainty of the risk estimates can occur if the community average health outcomes display spatial autocorrelation. That is, health responses for communities close together are more similar than responses for communities farther apart, thus invalidating the use of standard statistical models such as linear, logistic or hazard models. Autocorrelation in the residuals of these models could be due to missing or systematically mis-measured risk factors that are spatially autocorrelated. Failure to account for spatial autocorrelation can yield downward biased estimates of uncertainty in the community-level risk factors and may suggest uncomplete control for potentially confounding community-level factors with the variables of primary interest, such as air pollution (Miron 8 ) . We present a regression model in which the residual community health responses are characterized by community-based stochastic variables called "random effects", after controlling for individual and community-level risk factors. The variance of the random effects represents the residual variation

79

in response between communities. Broader spatial trends in residual health outcomes are modeled by non-parametric smoothers of location in the deterministic component of the model. This approach is analogous to that used in time series studies in which temporal trends in health series are jointly modeled with air pollution (Cakmak et al. 2 ) . Our modeling framework can accommodate continuous, binary, and time to event health data. We present an approach to statistical inference that can accommodate very large datasets typically encountered in this area. Our methods are illustrated using data obtained from the American Cancer Society (ACS) study of particulate air pollution and health (Pope III et al.g). We examined the association between the prevalence of heart disease and ambient particulate sulfate concentrations in 144 metropolitan areas in the United States. 2

Spatial Model

Three basic regression models are considred here: linear, logistic, and time to event. First, we start with time to event models. Suppose that the cohort of interest is stratified on the basis of one or more relavant covariates. Let the instanteneous probability of event at time t, or hazard fuction, for a individual i residing in area s and a member of stratum m is given by h™(t). We propose a space-time model to relate spatial risk factors to longevity. The hazard /ijm(£) for our model is determined by

jlm^e{(^)+pTxr(t)+0Tz(s)+V(s)}

^

where /i™(£) is the baseline hazard function for the mth strata, ((s) is the two-dimensional term to account for residual spatial variability, p is a vector of unknown regression coefficients linking individual risk factors to the hazard function, and /? is a vector of unknown regression coefficients linking the spatial level risk factors to the hazard function. Covariate information modulates the baseline hazard function with the regression parameters p and P representing the logarithm of the relative risk of death per unit change in the individual and spatial covariates, respectively. The spatial random effects, T](s), or frailties, are shared by all individuals in area s. These random effects reflect the difference between the observed hazard function and the hazard function predicted from a statistical model. We assume that the spatial process has zero expectation, variance , 0 > 0 and correlation matrix Ct with dimension S. Second, we consider logistic model. Let us assume that the response probability is given by n. The odds of positive response {j~) for a individual

80

residing in area s and a member of stratum m is defined by T T e{C(s)+p xr(t)+P z(s)+v(s)}

(2)

where C(s) is the two-dimensional term to account for residual spatial variability, p is a vector of unknown regression coefficients linking individual risk factors to the odds of positive response, /3 is a vector of unknown regression coefficients linking the spatial level risk factors to odds of positive response, and the spatial random effects process, 7j(s), are shared by all individuals in area s. Third, we consider linear model. The model assumes that dependence of health responses on covariates occours through linear combination: as)+pTXr(t)+0TZ(s)+r1(s)

(3)

where C(s) is the two-dimensional term to account for residual spatial variability, p is a vector of unknown regression coefficients linking individual risk factors to the health responses, (3 is a vector of unknown regression coefficients linking the spatial level risk factors to the health responses, and the spatial random effects process, r](s), are shared by all individuals in area s. 3

Statistical Estimation

We divide the estimation procedure into two stage. In "Stage One", we work in time domain. Health outcome data is modeled by covariates at the individual level and indicator functions for each community. Community-level covariates, such as air pollution, are not included at this stage. For our hazard model (1), for example, in the time domain we consider h™e&s.^5(s)I(>)+PT*r(t)}

(4)

where I(s),s = 1, ...,S — 1 are indicator variables taking the value 1 if the subject resides in area s and zero otherwise. One area (S) is (arbitrarily) assigned as a reference. The unknown parameters represent the logarithm of the relative risk of death for those subjects living in area s compared to those subjects in the reference area 5, after controlling for the individual risk factors Xi^(t). In the time domain the corresponding equations for linear and logistic regression models are straightforward. Estimates of the community-specific health responses are determined using standard computer software for linear, logistic and Cox proportional hazard survival models. We used SAS 10 procedures because they can accommodate very large sample sizes. For linear and logistic regression models, we do not specify an intercept term. The estimates of the indicator functions can be

81

interpreted as the average response in a specific community after adjusting for individual-level covariates. Output from stage one is the community-specific adjusted health responses denoted by {J(s),s = 1,...,S}, where s denotes a zero dimensional point in Cartesian (x,y) space representing the location of one of communities under study. Additional output from this stage is the variance-covariance matrix of the S(s), denoted by v, which describes the uncertainty in the estimates of the community-specific adjusted health response. The Cox proportional hazards survival model incorporates a baseline hazard function that is interpreted as the instantaneous probability of an event given all covariates in the model are assigned a zero value. In this case, indicator functions for community are defined with respect to a reference community. A limitation of this procedure is that the uncertainty of the estimate of the reference area is not defined. Because these values are based on comparisons with the same reference area, they are correlated. This correlation increases the estimated uncertainty in the location-specific log-relative risks {J(s)}. The induced correlation can be removed by methods developed by Easton et al. 4 . This procedure eliminates the covariance between the {<5(s)}, and defines an associated estimate of uncertainty to the assigned value of zero for S(S). In "Stage Two", we work in space domain. Estimates of community health responses are related to risk factors defined at the community level using a linear random effects regression model as follows: 6(s)=as)+PTZ(s)

+ r)(s) + e(s)

(5)

where e(s)is a random process with zero expectation, variance-covariance matrix v, independent of the spatial random effects process rj(s). C(s)is the two-dimensional trend term to account for residual spatial variability, and /? is a vector of unknown regression coefficients linking the vector of spatial level risk factors, Z(s), to the community-specific health responses. The spatial random effects, T](S) , are shared by all individuals in area s. These random effects reflect the difference between the observed and predicted values from our statistical model. We assume that the spatial process r](s) is stationary (i.e., expectation does not vary in space), has zero expectation, variance 0 > 0, and correlation matrix 0 with dimension S. The correlation of the random effects between two areas can be modeled by their distance apart or some other characteristic of their locations. Autocorrelation models, or correlation between the same response at different locations, typically assume that closer locations will have attribute values, in this case random effects, that are more similar than values in locations farther apart. Thus, these models are often characterized by functions that decrease monotonically with

82

distance (Matheron 7 ) . Here, <5(s), has expectation /u(s) = C(s)+0TZ(s) and variance-covariance matrix £ = OfJ + 1/ If the number of subjects and events in each community is large, as is assumed here, the {<$(s)} have approximately a multivariate normal distribution with mean vector /u and variance-covariance matrix £. The unknown parameter vector, f3 , and the variance of the random effects, 0 , can be estimated by maximum likelihood methods with the log-likelihood function minimized by Fishers Scoring algorithm (see Burnett et al. 1 ) . The elements of V are assumed to be known in this stage. The random effects will be non-stationary if there is a trend in space. Spatial trends can be accounted for by surface ((s). We consider non-parametric smoothed estimates of £ using the robust locally-weighted regression (LOESS) smoothers (Cleveland and Devlin 3 ) within the generalized additive model (GAM) framework (Hastie and Tibshirani 5 ) . The complexity of the surface is controlled by a "span" parameter which is the proportion of the data used for the local regression. The larger the span, the smoother the estimated surface. We have developed a simple method to judiciously select the appropriate span in the LOESS smoother so as to minimize the autocorrelation structure of the random effects. We do this by plotting the correlation of the standardized residuals versus the distance between communities. The span is decreased until there is no apparent association with distance. The GAM procedure in S-Plus requires that the data are uncorrelated. Within our iterative estimation procedure, we run the GAM with a weight function as the inverse of diagonal elements of S. The GAM regression model consists of a LOESS surface of (x,y) and spatial or community-level covariates, Z(s). We then capture the marginal values of this surface in the x and y dimension. These values are used as two new covariates in the random effects linear regression model, in which estimates of the covariate parameters, /?, and the dispersion parameters, 0 , are obtained by maximum likelihood methods. We iterate between the GAM step and the maximum likelihood step until little relative change in consecutive estimates of 0 is observed (in this case

< io- 4 ). 4

The American Cancer Society Study

Volunteers of the ACS enrolled over 1.2 million people in September of 1982 throughout the United States. Information on history of disease, longevity, and demographic characteristics were obtained. We obtained information on particulate sulfate levels from the Aerometric Information Retrieval System

83

(AIRS) and the Inhalable Particle Network (IPN) for 1980 and 1981 for 144 Metropolitan Statistical Areas (MSAs) in which ACS subjects were enrolled. Sulfates are secondarily formed particulate aerosols originating from sulfur dioxide emissions and are a major component of fine particulate matter. The sulfate data from AIRS was collected using glass fiber filters, which react in the presence of sulfur dioxide and artifactually innate the sulfate concentration. The sulfate data obtained from the IPN used teflon niters which are not subject to this artifact problem. Both monitoring networks were operating in 41 MSAs. We calibrated the AIRS sulfate data to the IPN sulfate data using six linear regression models with separate calibrations for three regions of the county and two time periods [April-September and October to March] 6 . Estimates of exposure were obtained by averaging all available sulfate data from all monitors located in a MSA for the years 1980 and 1981, inclusive. The mean concentration of sulfate particles adjusted for the sulfur dioxide artifact across all 144 cities was 6.4 ng/m3 , with a minimum value of 1.4 fig/m3, an interquartile range of 4.2 fig/m3, and a maximum value of 15.6 /j.g/m3.

5

Results

We examined the association between concentrations of sulfate particles and the prevalence of doctor diagnosed heart disease (8.6%). The mean age at enrollment for the 540,679 subjects was 56.6 years, 5% of subjects were younger than 40 years, 5% were older than 75 years, 56.6% of subjects were women, and 94% were white. There were 3,755 subjects per community on average, with a mean of 323 subjects with heart disease per community. BeaumontPort Arthur, Texas provided the least number of subjects (61) and the fewest with heart disease (7), while the most subjects were recruited in Los Angeles (23,151) with the highest number with heart disease (1,970). The first step in the analysis was to use the logistic regression procedure in SAS to identify all relevant individual risk factors that were associated with heart disease. This model also included indicator functions for the 144 communities to account for any extra between community variation not account for by the individual-level covariates. Thirty-seven risk factors were selected including variables representing age (a cubic polynomial to account for potential non-linear association with disease), gender, age-gender interaction, race (white versus other), quadratic polynomial of the number of cigarettes smoked and years smoking for current and former smokers, age starting smoking for current and former smokers, consumption of beer, wine, and liquor, quadratic polynomial of body mass index (square of height divided by weight), edu-

84

cational attainment (less than high school, high school, or more than high school), martial status (single, married, other), passive exposure to tobacco smoke, and regular exposure to some air toxics in work or daily life (asbestos, chemicals/acids/solvents, coal or stone dusts, coal tar/pitch/asphalt, diesel engine exhaust, or formaldehyde). In the next step, we visualized the spatial association between the prevalence of heart disease and sulfate particles using our stage two linear random effects regression model. Here, we regressed the {(5(s)} onto the (x,y) geographic coordinates defined by longitude and latitude of the 144 MSAs with a non-parametric smoothed spatial surface £ (Figure 1, right panel), excluding spatial covariate information such as air pollution (i.e. Z(s)=0 ). We used the GAM procedure in S-Plus 12 assuming no autocorrelation (i.e. U = 0) and a diagonal form for V. [A diagonal covariance matrix is required to use the GAM in S-Plus 12.] We use latitude and longitude for this visualization step because these coordinate definitions are more easily interpretable than the Cartesian (x,y) coordinate specification. We use the Cartesian coordinates in all other formal statistical analyses as the examination of spatial autocorrelation usually relies on Euclidian rather than angular distance measures. This procedure produced a three-dimensional surface of {^(s)} after adjusting for all individual level risk factors. We found that adjusted prevalence of heart disease was elevated in the Ohio Valley region south of Lake Erie and in the southeast, diminished in the west and south, and moderately elevated in the mountain states. We also spatially modeled concentrations of sulfate particles using a GAM assuming these values were uncorrelated (i.e. E = a21, where a2 is the residual variance and / is the identity matrix). Modeled sulfate values centered by their mean concentration are portrayed in left panel of Figure 1. A corresponding elevation in concentrations of sulfate particles exits in the Ohio Valley region and the southeast, with much lower concentrations in the west. Heart disease was elevated in the mountain states, a pattern not observed for sulfates. This visualization suggests a positive association between the two surfaces. We then fit our spatial linear random effects model with no spatial predictors, no location surface, assuming that the data are uncorrelated (S = 0 / -I- V) and determined the standardized residuals from this model. The association between the autocorrelation of these standardized residuals and distance is graphically presented in Figure 2 (panel a) using the correlogram function in the Spatial Module of S-Plus u . Correlograms measure autocorrelation as a function of distance between the observations within binned distance groups. This procedure generates average autocorrelations

85

Figure 1. Non-parametric smoothed surface of the prevalence of heart disease by latitude and longitude, adjusted for individual level covariates in American Cancer Society Study with smoothing parameter of 20 percent (left panel). Non-parametric smoothed surface of particulate sulfate concentrations by latitude and longitude with a smoothing parameter of 20 percent (right panel). Note, z-axis represents residuals from generalized additive model.

at 20 roughly equal distances. Autocorrelation peaks at a value of 0.45 for near communities, and is positive but declining for distances under 500kms.. A cyclic pattern is evident in the autocorrelations for distances greater than 500kms.. This pattern could be due to the several regions of elevated prevalence of heart disease (see Figure 1, right panel). The association between autocorrelation and distance is reduced slightly by the inclusion of sulfates (Figure 2, panel b). We also examined the autocorrelation structure in the standardized residuals for models with a LOESS location surface with spans of 80, 60, 40, and 20% (Figure 2, Panels c-f respectively). A span of 20% was required to eliminate the association between the standardized residuals and distance. The sensitivity of the sulfate effect, /3, and the random effects variance, 0 , to the model specification is given in Table 1. The sulfate effect based

86 b) Sulfate

a) No Spatial Predictors

O O

500

1000

Q_

1500

1000

1500

2000

distance (km)

distance (km)

c) Sulfate + Location (span=80%)

d) Sulfate + Location (span=60%)

o o o 500

1000

1500

2000

1000

1500

2000

distance (km)

distance (km)

e) Sulfate + Location (span=40%)

f) Sulfate + Location (span=20%)

f a o u 6 o « 500

° ° o u

< > " . . „ « »

1000 distance (km)

1500

2000

500

1000

1500

2000

distance (km)

Figure 2. Correlation of standardized residuals by distance between locations for spatial random effects model under six model specifications: a) no spatial predictors and no location surface b) sulfates as the spatial covariate with no location surface, c-f) model includes sulfate and a location surface estimated with a LOESS span of 80-20% respectively. Horizontal line indicates zero values.

on the standard logistic regression model ( Model 1) assuming subjects are independent 0 = 0.0100) was similar to the corresponding estimate based on our "Two-Stage" spatial random effects model ( Model 2) (/? = 0.0103) under the identical assumption of independence (i.e. 0 = 0 ) , indicating that the 2stage estimation approach gave similar results to the logistic regression model for this example. This was most likely due to the large number of subjects and cases of heart disease per MSA. The estimates of the standard errors of the sulfate effect for the two approaches were also similar (Table 1). There existed strong statistical evidence to support the assumption of

87 Table 1. Sulfate Effects and random effects variance by model type and span of LOESS smoother of location surface. Span(%)

Sulfate Effect(/3) (st. error)

Relative Risk* (95% Conf. Int.)

Random Effects Variance(0))

Logistic (1)

NA

0.0100 (0.0018)

1.043 (1.027, 1.058)

NA

Spatial (2)

NA

0.0103 (0.0019)

1.044 (1.028, 1.060)

0

Spatial (3)

NA

0.0123 (0.0033)

1.053 (1.026, 1.080)

0.0076 0.00041

Spatial (4)

80

0.0132 (0.0036)

1.057 (1.025, 1.089)

0.0062 0.00041

Spatial (5)

60

0.0132 (0.0035)

1.057 (1.026, 1.089)

0.0055 0.00015

Spatial (6)

40

0.0131 (0.0032)

1.057 (1.028, 1.085)

0.0043 0.00012

Model Type (Model No.)

Spatial 20 1.047 0.0110 (1.02, 1.07) (0.0031) (7) *: Relative risk evaluated at interquartile range of sulfate levels (4.2 NA: not applicable

0.0022 0.00025 ng/m3)

additional variation in the adjusted prevalence of heart disease between communities (i.e., 0 > 0) based on the likelihood ratio test comparing Models 2 and 3 (p < 0.0001). Increasing the complexity of the location surface (i.e., decreasing the span) was also justified in terms of improving the model's fit to the data (p < 0.0001) (likelihood ratio tests comparing Model 3 to Models 4 to 7 respectively). The inclusion of a random effect for location ( Model 3) increased the estimate of the sulfate effect 0 — 0.0123) but almost doubled the estimated standard error (0.0033) compared to the error obtained from a model assuming independence among subjects (i.e., 0.0019). This suggests that there was more variation in heart disease between communities ( 0 = 0.0076) than can be fully explained by the within community estimation error,V, and sulfate particles. The sulfate effect estimate was insensitive to inclusion of location surfaces (Models 3-7). The unexplained between community variation, however, was much lower for Model 7 ( 0 = 0.0022) compared to Model 3 in which no surface

88

"'ini

2

mi

[I inn HI

4

IMII[

6

urn i

I

i

8

ninii

10

—LU

12

, U

14

L^.

16

sulfate

Figure 3. Cubic-spline representation of the association between particulate sulfate concentrations (fig/m3) and prevalence of heart disease adjusted for individual level covariates and a smoothed representation of location (LOESS span of 20 percent). Dashed lines represent 95 percent confidence intervals, with tick marks indicating sulfate values.

modeling was undertaken ( 0 = 0.0076), reflecting the ability of the location surface to explain between community variation not accounted for by sulfate particles. The form of the relation between sulfate concentrations and the prevalence of heart disease is illustrated in Figure 3. Here, we used a spline function representation of sulfates in the GAM after adjusting for a LOESS surface of heart disease using a span of 20 percent. There was little statistical evidence for a departure from a linear association (p=0.4429).

89 6

Discussion

We have identified two related issues about the analyses and interpretation of studies linking spatial variation in ambient air pollution and health. First, if the assumption of statistical independence is not valid, the uncertainty in the estimates of effect may be understated. Second, autocorrelation in these residuals may suggest that there exists missing or systematically mis-measured risk factors that may be correlated with air pollution and consequently could confound the pollution-health association. On the first issue, our model provides more accurate estimates of the uncertainty of estimates of effect. Based on the analysis of the ACS data, while our model gave similar sulfate-heart disease estimates as the standard logistic model, the standard errors of these estimates were higher (Table 1). With regard to the second issue, we have observed a pattern of spatial autocorrelation in the prevalence of heart disease that cannot be fully explained by ambient particulate sulfate concentrations, even after controlling for a host of risk factors measured at the individual level. Inclusion of a location surface eliminated this spatial correlation pattern and slightly reduced the uncertainty in the sulfate effect estimate from 0.0030 for a LOESS location surface estimated using a 80% span (Model 4) to 0.0029 for a span of 20% (Model 7). However, the sulfate effect was insensitive to adjustment for spatial trends in heart disease suggesting that the association between heart disease and sulfate pollution was strong enough to effectively complete with location in predicting the prevalence of heart disease. We have extended our modeling approach beyond that reported in our reanalysis 6 . Spatial autocorrelation was modeled by including regional indicator variables into the random effects model. This procedure was ad hoc in that there is no unique method for defining regions. Our new approach of using location surface models allows the data to determine the form and extent of the spatial adjustment. We also removed spatial autocorrelation by pre-filtering both the S(s) and the sulfate data for spatial trends using a spatial moving average function. The radius of data inclusion for the moving average term was based on the distance beyond which no evidence of spatial autocorrelation was graphically apparent. Furthermore, the number of communities comprising the moving spatial average varies in space, yielding spatially filtered estimates with varying uncertainty. Finally, all evidence of associations between air pollution and mortality at the regional scale are removed using the pre-filtering approach. Our new modeling method allows air pollution to compete with location to predict health responses. Evidence of regional scale associations between health responses and air pollution will be

90 captured with this new approach. Our spatial models can also be used to assist with the selection of appropriate spatial or community-level risk factors. The visualization stage suggested that the upper midwest had elevated prevalence of heart disease and this pattern was not observed in the sulfate data. This would indicate the need for including additional risk factors that cluster in this region. By repeating the visualization and spatial dependence analysis after adding new variables, our method can help to identify variables most likely to explain significance between-community variation. References 1. R. T. Burnett, W. H. Ross and D. Krewski, Environmetrics 6, 85 (1995). 2. S. Cakmak, R. Burnett and D. Krewski, J. Expos. Anal. Environ. Epidemiol 8, 129 (1998). 3. W. S. Cleveland and S. J. Devlin, J. Am. Statist. Assoc. 74, 829 (1988). 4. D. F. Easton, J. Peto and G. A. G. Babiker, Statist, in Med. 10, 1025 (1991). 5. T. Hastie and R.Tibshirani, Generalized Additive Models, (Chapman and Hall, London, 1990). 6. Health Effects Institute, Reanalysis of the Harvard Six Cities Study and the American Cancer Society Study of Particulate Air Pollution and Mortality: A Special Report of the Institute's Particle Epidemiology Reanalysis Project, (Health Effects Institute, Cambridge MA,2000). 7. G. Matheron, Eco. Geology 58, 1246 (1963). 8. J. Miron, In Spatial Statistics and Models, eds. G.L. Gaile and C.J. Willmott (D. Reidel Publishing Company, Boston, 1984). 9. C. A. Pope III, M. J. Thun, M. M. Namboodiri, D. W. Dockery, J. S. Evans, F. E. Speizer and C. W.Heath, Am. J. Respir. & Crit. Care Med. 151, 669 (1995). 10. SAS/STAT Software, Changes and Enhancements through Release 6.12, (SAS Institute Inc., Cary, NC, 1997). 11. S+SpatialStats, User's manual for Windows and UNIX, (MathSoft, Data Analysis Products Division, Seattle, WA, 1997). 12. S-PLUS, Programmer's Guide, (MathSoft, Data Analysis Products Division, Seattle, WA, 2000). 13. J. H. Ware and D. O. Stram, Can. J. Statist. 16, 5 (1988).

O N T H E R O B U S T N E S S OF RELATIVE S U R P R I S E I N F E R E N C E S TO T H E CHOICE OF P R I O R M I C H A E L EVANS Department

of Statistics, University of Toronto, Toronto, E-mail: [email protected]

Department

of Statistics, University of Toronto, Toronto, E-mail: tianliQutstat.utoronto.ca

Ontario

M5S

3G3

Ontario

M5S

3G3

TIANLI ZOU

Relative surprise inferences are derived in the context of a sampling model for the data and a prior model for the parameter by examining how beliefs change from a priori to a posteriori. Relative surprise inferences are often different than those that arise from frequency or some standard a posteriori Bayesian methods. Given that relative surprise inferences are determined by how the data has caused beliefs to change we would expect that these inferences are more robust to the choice of prior than inferences that are dependent solely on the posterior. This paper is concerned with demonstrating this and quantifying the extent to which robustness is enhanced by taking the relative surprise approach. The connection with Bayes factors is also explored.

1

Introduction

Suppose that we have model {/# : 6 € fi} for data x € X, where each f$ is a density with respect to support measure /x, and a prior density n with respect to support measure v for 6. Suppose further that our interest is in making inferences about the marginal parameter r = T (6) where T : ft —> T. Conditional probability (Bayes Theorem) then prescribes that inferences about T should be based on the conditional distribution of r given x; i.e. the posterior distribution of r. We will denote the prior density of r with respect to some support measure vj- on T by 7r-f and the posterior density of T with respect to vj- by 7TT (• | x). Throughout this paper all priors are proper. The actual choice of the densities is important for what follows and this cannot be done arbitrarily; e.g. changing the value of a density on a set of measure zero. For the purposes of this paper we will always take our densities to be limits. So, for example, if P is a probability measure on a space S, absolutely continuous with respect to support measure A, then the density at s will be taken to be equal to dP_ ,g) d\

=

P{An) lim n-too A (An)

91

92 where An C S is a sequence of sets that converge "nicely" to s. That we can define densities in this way in very general situations follows from definitions found, for example, in Rudin 8 and, as can be seen from Evans, Eraser and Monette 2 , this does not restrict our discussion in any meaningful way. Further this restriction on the definition of densities establishes an important connection between the relative surprise approach to inference and the use of Bayes factors as shown at the end of Section 2. In Section 2 we review some discussion found in Evans : concerning the use of surprise and relative surprise for deriving inferences. In addition we establish a relationship between the relative surprise approach to inference and the use of Bayes factors and consider an important example whose robustness properties are subsequently analyzed. Inferences based on what is called the observed surprise are seen to correspond to some standard Bayesian inferences. These inferences suffer from a lack of invariance; i.e. the inferences depend intrinsically on the parameterization chosen for the model. There may be reasons in a particular application why a parameterization must be fixed; e.g. a loss function is prescribed, but, in general, it seems reasonable to ask that our inferences not be parameterization dependent. Relative surprise inference, as defined in Evans 1, is one example of inference that is independent of the parameterization. This type of inference is also seen to have a very natural interpretation as it is based on how the observed data has changed our beliefs from a priori to a posteriori. Given that relative surprise inferences are more data driven that some standard Bayesian inferences one might expect that they are more robust to the selection of the prior. This is the topic of Section 3 and is the main content of the paper. We develop the relevant mathematical theorems for characterizing the local robustness properties of relative surprise and surprise inferences and compare these. Although the arguments are somewhat subtle the results confirm our suspicions that relative surprise inferences exhibit superior robustness properties when considering sensitivity to the choice of prior. Our approach to studying the robustness properties of our inferences is similar to that taken in Gustafson and Wasserman 6 and Gustafson 4 ' 5 . One key difference, however, is that these authors are concerned with developing robustness diagnostics for applications of Bayesian inference; i.e. determining whether or not a particular model, prior and data combination is robust. While such diagnostics are highly relevant in applications, however, this is not our concern here. We simply want to compare the robustness properties of two different approaches to inference. In Section 4 we draw some general conclusions and discuss future research on this topic.

93 2

Surprise and Relative Surprise

While conditional probability prescribes that inferences should be based on the posterior distribution it says nothing about what form these inferences should take. For example, if we wish to quote an estimate of r then perhaps we might think of using the posterior mean as natural choice. Of course this may not make sense in general as r need not take its values in a Euclidean space, or the relevant expectation may not exist or the posterior distribution may be strongly skewed or multimodal which could render its mean an inappropriate choice as a representative value. One way to prescribe an inference in such a situation is to specify a loss function and to use a Bayes rule; e.g. posterior expectations are the optimal choice when using squared error loss. Such an approach places a premium on minimizing expected losses over any other considerations. Another approach is to look for a principle or basic idea that leads necessarily to the form of the inference much as the principle of conditional probability leads to the use of the posterior. If the principle, or basic idea, is compelling then we feel confident that the inferences derived make sense. Of course the assessment of the principle involves examining many applications to see whether or not its application leads to what are generally considered to be sensible inferences. One principle, or basic idea, is that of surprise. In essence we want to measure whether or not a specified value TQ € T is surprising in light of the data x. While there are a number of different ways of measuring surprise one fairly natural method for this is given by the observed surprise (OS) n ( 7 r T ( T ( 0 ) \x)>*r{TQ\x)\x);

(2)

namely the posterior probability that the posterior density at r = T (9) is greater than at the specified value TQ. If (2) is near 1 then To is in a region where the posterior density is relatively small and r 0 is then said to be surprising in light of the data. As an example consider the following. Example 1. Hypothesis testing. Suppose that T = IH0 (the indicator function for the set H0) where H0 C fi is such that II (Ho) > 0 and r 0 = 1. So in such a case (2) provides an assessment of whether or not the hypothesis Ho is surprising in light of the data. Taking 7rT (11 x) — II (H0 | x) and 7TT (01 x) = II (H£ | a;) we have that

94

the observed surprise equals II (TTT (T (9) \X)>TTT

_/0

{T0

\X)\X)

]SU{H0\x)>U{HS\x) c

(

c

, {

~ \ n (#0 | re) if n (H0 i x) < n (#0 j x) ' and we have evidence against Ho when II (HQ | X) is large. It is easy to see that (3) is a Bayes rule when using 0-1 loss so that the OS approach agrees with the standard Bayesian decision-theoretic answer for this problem. Sometimes it is argued that posterior probabilities are to be preferred to P-values in assessing a hypothesis but (3) makes it clear that assessing hypotheses using posterior probabilities can also be thought of as falling within the P-value approach. The virtue of a P-value approach as exemplified by (2) is that it does not require II (Ho) > 0 to provide a measure of surprise, although the assessment is different than simply looking at T = i # 0 , so that continuous priors can be used. This has some consequences for the avoidance of Lindley's paradox, as discussed in Evans 1, but this issue does not concern us here. If we want to estimate r then Good's principle of least surprise naturally leads us to select the least surprising value of To and this is a value which minimizes (2). It is obvious then that the least surprise estimator is given by a mode of the posterior density iry (• | x). Further if we wanted to construct a region containing the true value of r then perhaps a natural choice is to specify 7 € [0,1] and use the 7-surprise region {T0 I n (TTT (T (0) \x)>nr

(r 0 | x) \ x) < 7} ;

i.e. the set of values for To that are not surprising at the level 7. It is immediate that such a region corresponds to a highest posterior density (HPD) region for r. The above shows that measuring surprise via (2) leads to some standard Bayesian inferences and that the derivation of these does not require the specification of a loss function. It has been pointed out previously in Evans 1 , and it may be obvious to the reader, that there is a fundamental difficulty with inferences derived via (2). In particular these inferences will depend upon how we specify the posterior density 7TT (• | x), or equivalently, how we specify the support measure v-r- Different choices will lead to very different values for (2), very different estimators and very different 7-surprise regions. In general there does not seem to be an argument that leads us to a canonical choice of support measure. So surprise as a justification for these inferences seems untenable and perhaps we are even lead to doubt the validity of such inferences even though posterior modes and HPD regions are commonly used.

95 As a solution to this problem Evans l proposed to base the derivation of inferences instead on the observed relative surprise (ORS) given by JJ | 7TT (Y (0) | x) ^

nr(T(6))

7Tr(T0\x)

(4)

7TT (To)

Here 7TT (TQ I x)

(5)

7TT ( r 0 )

is measuring the change in belief in TQ from a priori to a posteriori and (4) is the posterior probability of a change in belief larger than that observed at T 0 . Again if this number is near 1 we have that To is a surprising value. In essence a value To is surprising when the data has lead to a bigger increase in belief at other values of r = T (6) compared to the increase (or decrease) at T0. A least relative surprise estimate is then obtained by choosing a value of r 0 that maximizes (4) or equivalently maximizes (5). As above we can also obtain 7-relative surprise regions. A virtue of all the relative surprise inferences is that they are not dependent on the choice of the prior or posterior densities or equivalently the choice of the support measure vj-- These inferences are invariant under smooth reparameterizations. As shown in Evans 1 these inferences can also be quite different than standard Bayesian inferences although often they are similar. We consider the context of Example 1. E x a m p l e 2. Hypothesis testing (Example 1 continued). We consider again the situation where T = IH0 , II (Ho) > 0 and we take 7TT (1) = II (Ho) and 7Tx (0) = II (HQ) then the observed relative surprise is given by n

J' 7TT ( T (
>

7TT ( T 0

I a;)

71"T ( r 0 )

if BFHo > 1 Ho
-{n {H§\x)i£BF

(6)

where

BF,Ho

U(H0\x) l-U(H0\x)fl

n {H0

n(H0)

(7)

is the Bayes factor in favor of Ho- So we get something similar to what is commonly recommended in this context. We will see, however, that the robustness properties of (6) are very different than those of (3). Further (4) has the virtue of providing a measure of surprise even when II (Ho) = 0 and again, see Evans 1 , this has implications for the avoidance of Lindley's paradox.

96

Sometimes just the Bayes factor is recommended in hypothesis testing with small values of (7) treated as evidence against Ho- The interpretation of the value a Bayes factors takes is somewhat ambiguous, however, although various calibrations have been suggested; see, for example Kass and Raferty 7 . Perhaps the most direct way of calibrating a Bayes factor is to proceed as we have with the OS and ORS computations and compute the posterior probability of obtaining a value larger than (7) with large values of this probability indicating that Ho is indeed surprising. For the context of Example 1 we obtain

= {0u

n{BFr(9)>BFHo\x)

if BFHo > 1 (H£ | x) if BFHo < 1 ;

i.e. this measure of surprise agrees exactly with the ORS. So in a sense we can think of the ORS as a generalization of the Bayes factor when we choose to calibrate the value of a Bayes factor using the tail probability. Suppose more generally we have that T = { r i , . . . , r*} with prior probabilities 7Ti,..., n-* respectively, and we want to assess the plausibility of the value To. Then U({T}\X) T

n({r}) i-u({T}\x)'i-n({T})

and possibly we could assess the surprise at To by computing n (BFr(e)

> BFT0 | re)

but when k > 3 this is generally not equal to the ORS at To- Effectively this approach is saying that the change in belief in r from a priori to a posteriori should be measured by BFT and this seems somewhat less direct than using 7TT (r | x) /-7TT (T) as in the ORS. Further it is not at all clear how to extend this approach to the continuous case. Notice, however, that when II (An) > 0 for all n, and the An converge nicely to {r} with II ({r}) = 0, then _ n (An I X) IVT (An) A

"

l-U(An\x)

II ( H ) '

/vr

l-U(An)

(An)

7TT (T I x) 7TT(T)

as n ->• oo. So in general it is very natural to think of the relative surprise approach as a modification and generalization of the Bayes factor approach. Of course the Bayes factor is restricted to hypothesis testing problems where the parameter space is partitioned into ft = H0UHQ with the prior probability of H0 satisfying 0 < II (H0) < 1. The relative surprise inferences are more general in their application than this, however, both with respect to hypothesis testing, and estimation, prediction and model checking problems; see Evans

97 3

Robustness to the Prior Under Linear Perturbations

In an obvious sense relative surprise inferences are more data driven than inferences based on surprise. Observe also that when T (6) = 9 then (4) equals n {fe{x)>

fe0{x)\x);

i.e. the observed relative surprise is the posterior probability that the likelihood at 8 is greater than the likelihood at 9$. In particular, a least relative surprise estimate of 6 is given by a maximum likelihood estimate while relative surprise regions for the full parameter 9 correspond to likelihood regions and only depend on the prior through their posterior probability contents. So in a general sense we expect relative surprise inferences to be less dependent on the prior than other Bayesian inferences. It is the purpose of this section to examine this issue more closely and in particular to examine the marginal parameter case; i.e. T (8) ^ 9. Our interest then is in how the inferences vary as we perturb the prior II. Basically we will say one inference method is more robust than another if the rate of change in the inference is smaller at II for the first inference method than for the second. For example, it is clear immediately that when our goal is to estimate 9 then a least relative surprise estimator is more robust to the choice of prior than any other Bayesian inference because it only depends on the likelihood function. Actually this example is somewhat unusual as even the estimation problem becomes more difficult as soon as interest is in a marginal parameter r. In that case a LRSE will depend on the prior and the extent of the dependence is not obvious. Still the result for the full parameter leaves us hopeful that something similar can be obtained for a marginal parameter as well. Computational and mathematical issues prevent us from arbitrarily choosing the directions in which we perturb the prior so that definitive comparisons; i.e. results that hold for all directions, seem impossible to achieve at this point. Still we can obtain results for several choices of directions and the results give us considerable insight into the relative robustness of the various inferences. Further, since all the inferences under consideration are based on the observed surprise (2) or observed relative surprise (4), we will compare the robustness of these quantities. Recall that these quantities can be used for assessing whether or not ro is a plausible value for r. Perhaps the most commonly used directions are the contamination directions given by prior probability measures of the form

IP = (1 - e) n + eQ

98 where e > 0 and Q is a probability measure on Q. The prior density of 9 with respect to v is then given by 7re = (1 - e) 7T + eq

when Q is also absolutely continuous with respect to v. Actually, because our interest is in the marginal parameter r, we want to consider only perturbations to the marginal prior distribution so that the conditional distribution of the full parameter given the marginal remains fixed. This kind of perturbation is easily expressed in terms of densities as Tre (9) = (1 - e) 7TT (r)

TT

(0 I r) + eqr (r)

IT

(9 | r)

= [(l-e)7rT(T)+«fr(T)]7r(0|T)

(8)

where TT(6\T) is the conditional density of 9 given T (8) = T with respect to support measure vT on {9 : T (9) = r } . Then the marginal posterior density, when the prior density is given by (8), is Try (T I x) = (1 - ex) TTT (r | x) + exqr (r | x) where £x

em

=

« (*) (1 - e) m^x) + emq (x)'

(9)

™*(x) = /" / 6 (K)ir(e)u(de) is the prior predictive density for the data when the prior is 7r with a similar definition for mq (x), 7Tr (T I x) —

Qy

(T

I x)

TT (r)/,*(*)

mn (x) 9r(r)f;(x) mq(x)

and we write the conditional density with respect to /j, of x given that T (9) = T as JT-HT}

Now letting G denote the posterior distribution function of ny (r | a;) when T ~ ITx (• | x) then the following result, as proved in Evans and Zou 3 , yields expressions for the upper and lower Gateaux derivatives of the observed surprise.

99 Theorem 1. If the posterior distribution of ny (T\X) is continuous when T ~ n T (• | a;) or when T ~ Qy (• | x), G is differentiable at -Ky (TO \ x) with derivative g (ny (r 0 | x)) and <7T (• | x) is continuous and bounded, then the lower Gateaux derivative (and the upper Gateaux derivative) of the observed surprise at r 0 , under linear perturbations of the marginal parameter r = T (0), takes the form mq (x) {qr (r» \x)-qr (T 0 | X)} g (ny (r 0 | a;)) + mn(x) mq{x) | <3T [TTT (r I x) > TTT (r 0 | a;) | x] - 1 m^x) \ n T [TTT (r | a;) > TTT (T 0 | X) \ x] J

._

for some T* Of course when the Gateaux derivative of the observed surprise exists it takes the form given in (10) as well. The expression (10) is not very useful when we need to compute the Gateaux derivative as r* is unspecified. In specific examples we are still required to evaluate the limit corresponding to the first part of (10). We will see, however, that (10) is of great value when comparing the robustness of the OS with the robustness of the ORS. We note that the second term in (10) is always bounded above by S

-^jJlM mn[x)

{l _ n T [TTT (r | x) > TTT (TO I x) I a:]} .

The following result, as proved in Evans and Zou 3 , suggests that the first term in (10) can be arbitrarily large in absolute value. Lemma 2. If the density function g of Try (r | a;) is continuous from the left at 7Tx (TO I x), where To is the unique mode of 7TY Ola:), and 7TT (• | x) is continuously differentiable at TQ then g(Try (TQ \ x)) = oo. So Lemma 2 indicates that the first term in (10) could be infinite whenever To is a posterior mode. That this happens is confirmed by the following example. Example 3. Gateaux derivative of the OS can be infinite. Suppose that x ~ Bernoulli^ 1 / 2 ) where 8 G [0,1] is unknown, we take II to be the Beta(l/2,1) distribution and the support measure v is Lebesgue measure. Then the posterior density, when we observe x — 1, is given by 1; i.e. the posterior distribution is uniform. Suppose we take Q to be a Beta(3/2,1) distribution which leads to the posterior density 28 when we observe x — 1. Now we explicitly evaluate the first term in Gateaux derivative of the observed

100

surprise at 60. First note that II [IT (6 \ x) > it (90 | x) \x] = U [1 > 11 x] = 0 and (1 - ex) + 2ex6 > (1 - ex) + 2ex60 if and only if 0 > 60. Therefore

lim - ( n (1 - cx) + 26*19 > e-+0 e L

(1 - ex) + 2ex60

n

[TT (6 |

x) > n (d0 | x) | a;]

= lim - (1 - 60) = oo e->0 e

and we see that the first term in the Gateaux derivative for the observed surprise is infinite for every #oIn Evans and Zou 3 the following result is established for the Gateaux derivative of the observed relative surprise. Theorem 3. The Gateaux derivative of the observed relative surprise at To under linear perturbations of the marginal parameter r = T (6) is given by

^ { Q T [ / ; ( * ) > / ; ( * ) ! x ] - n r [/;(*)>/;»|*]}.

(n)

lil-n^X)

Note that this result implies that the Gateaux derivative of the observed relative surprise is always finite. In particular consider an application of this result to Example 3. Example 4. Gateaux derivative of the ORS in Example 3. We have that fg (x) = 61/2 and therefore Q [/; (x) > f*T0 (x)\ x] =Qr[6>

do I x]=

f

29d0 =

l-el

n[/;(x) > /;o(x)\ x] = u[e>e0\ x] = [ de = i-e0 J00

while mn(l) = 1/2 and mq (1) = 3/4. Therefore the Gateaux derivative of the observed relative surprise at #o is given by, using (11), | (1 - 0g - 1 - (?o) = |«o (1 -
101

of Q measures that converge to a measure degenerate at a LRSE then (11) converges to

s

-^im{1_Ur[rAx)>f:Ax)\x]}

(12)

and this is bounded whenever sup r / * (x) < oo. It clear then that (12) is the supremum of (11), when the supremum is taken over Q. We now establish a result that formally confirms our intuition, at least in the continuous case, that inferences based on the ORS are more robust to the choice of prior than inferences based on the OS. For this we consider the supremum of the absolute values of the Gateaux derivatives over all possible absolutely continuous perturbation measures Q and all possible reparameterizations. By a reparameterization we mean a 1-1 bicontinuously differentiable map defined on T. Theorem 4. If r has an absolutely continuous prior distribution on an open subset of a fc-dimensional Euclidean space, then the supremum of the absolute value of (10) is greater than or equal to the supremum of the absolute value of (11) when the supremum is taken over all possible absolutely continuous measures Q and all possible reparameterizations. Proof: When ir is absolutely continuous on an open subset of Rk we can reparameterize the problem so that the prior distribution for r is uniform on [0,1] . For example, we could use the probability transform based on ir to do this. With this parameterization, if To is a LRSE then it is also a posterior mode so that (11) is 0 and so is the second term of (10). From Lemma 2, however, we have that the first term of (10) is either 0 or infinite in absolute value. So in the case that To is a LRSE the supremum, over all possible parameterizations, of the absolute value of (10) is always greater than or equal to the absolute value of (11). When ro is not a LRSE then reparameterizing so that r is uniform on [0,1] establishes that the second term of (10) equals (11). Further if we take a measure Q that assigns 0 probability to a neighborhood of To then
102

view of robustness one would want to ensure that such a selection did not lead to inferences that had terrible robustness at least when using some traditional Bayesian inferences. Of course this is not an issue when using relative surprise inferences as these are not dependent on the parameterization chosen. We interpret Theorem 4 as strong support for the increased robustness of the ORS over the OS. Theorem 4 is restricted to contexts where the prior is absolutely continuous. Discrete contexts seem much more difficult to analyze. We consider now, however, an important problem where the marginal prior is discrete. Example 5. Hypothesis testing (Examples 1 and 2 continued). Consider the problem discussed in Section 1 where we want to test the null hypothesis HQ and we have II (HQ) — no > 0. Therefore the posterior probability of HQ is given by V"O/H

n(H0\x)=

(X)

° J(1n o- _ TO) ^ f*m ^OIHO ( — ) + X

(x)'

Let Q be the measure degenerate at HQ. Under a small perturbation e of the prior probability the posterior probability becomes W(H0\x)~

( U - < ) *o+ 0/*„(*) ((1 - e) TTO + e) rHo (x) + (1 - c) (1 -

/ £ . (x)'

TTO)

Now suppose that II (H0 \x) > U (HQ \X) . Then, by (3), and the fact that IF (Ho \x) > IP (HQ | X) for all e small enough, the observed surprise is 0 when e is small enough and so the Gateaux derivative of the observed surprise is also 0. If n (H0 | x) = n (H§ | x) then 7r 0 /£ 0 (x) = (1 - TT0) / £ g (x) and so IT _

(H0\x)

((l-e)Tro + e) fHoW ((1 - e) no + e) f*Hg (x) + (1 - e) (1 - n0) f*H< (x) ni-xpU

I

( (l-e)7ro

+e )

TTO

j\(l-e)(l-no)j

f

(l-e)(l-*o)rHS(x)

]

\ ((1 - e) TTO + e) f*Ho (x) + (1 - c) (1 - 7T0) f*H, (x) j 1 /

e \ f 7^0 +

TTO V

1-e/

(l-e)(l-7r0)/£c(i)

1 ((l-e)7ro+e)/^0(x) + (l-e)(l-7ro)/^(a;)

103

(l-e)(l-7ro)/&.(s) > ( ( i - c ) *•„ + e ) fa (X) + (1 _ e ) (1 -

=

TTO)

fa, (x)

W(HZ\x)

and the Gateaux derivative is again 0 by (3). If U(H0\x) < II (#0C | x) then, since IP (H0 \x) < IF (#0C | x) for all e small enough, the Gateaux derivative of the observed surprise is given by I f

((1-<0*O+C)/HO(S)

~ i ™ 7 \ ((l- £ )7r 0 +e)/^ o ( a : )+(l- e )(l-7ro)/^(x) = - (*0fk

=

TO/£„(*)

7ro/£ 0 (z)+(l-7ro)/£ s

(X) + (1 - TTo) / £ « (*)) _ 2 (1 - TTo) f*Ho (x) ft- (X)

-—n(H0\x)Il(HZ\x) TTO

(1 - TTo) BFHo (l-7r0+7ro5F.ff0)

(13)

2-

Now observe that BFfj = jBF/fo for all e; i.e. the Bayes factor in favor of HQ is independent of e. This implies that the Gateaux derivative of the observed relative surprise is given by 0 when BFH0 > 1 and is given by (13) when BFHa < 1. It would seem then that the behavior of the two measures of surprise is very similar. Consider, however, the situation where BFH0 > 1, n (Ho \ x) < II (HQ I x) and U(H0) — 7r0 is very small. In this situation the Gateaux derivative of the ORS is 0 while for the OS (13) is approximately equal to BFH0 • So in situations where the data have lead to a large value of BFHo, but the posterior probability of Ho is still less than 1/2, the observed surprise is very sensitive to perturbations in the prior belief assigned to Ho. Now consider the situation where BFJJ0 < 1 while II (H0 \ x) > II (HQ | X) so that the Gateaux derivative of the OS is 0 and the Gateaux derivative of the ORS is (13). We have that (13) is bounded above by BFHo 1 - 7T0 + TT0BFHo

and, because BFJJ0 < 1, this is an increasing function of -KQ which goes to the value 1 as 7To -> 1. So we have shown that the worst case behavior of the Gateaux derivative of the ORS in this problem is given by the upper bound 1 while the worst case behavior of the Gateaux derivative of the OS, using the relation BFJJ0 =

104

/ H 0 ( X ) If HI (X) > i s s i v e n b y

sup ZkM. Based on this we see that the ORS has superior robustness properties to the typical Bayesian inference II (HQ \ x) for testing HQ versus HQ. As mentioned previously sometimes BFH0 is used in these contexts and this has the appearance of being fully robust because BFjjo = BFH0 for all e. Recall, however, the need to calibrate the value of a BFH0 through an equation such as (7). This leads to n(H5\x)=(l

+ Y^BFHo>)

.

Therefore a value of BFH0 = 20 when 7r0/ (1 - 7To) = 1 implies II (HQ \ x) = 1/21 = .048. On the other hand BFHo = 20 when TT0/ (1 - 7r0) = 1/10 implies II (HQ I x) = .5. The point here is that the interpretation of the value BFH0 does inherently depend on the values of the prior probability -KQ and so is not robust to this choice. The key concept in determining relative surprise inferences is the ORS and so it makes sense to concentrate on assessing its robustness properties. Ultimately, however, we should look at the robustness properties of the specific inference. With hypothesis testing this is the ORS but with estimation, for example, we need to look at the LRSE. We note that *?(T\X)=

-Ky (T)

f;(x)

(1 - e) m7r(x) + em, (a;)

for every e > 0 immediately implies the following result. Theorem 5. The LRSE of r is constant under perturbations that affect only the marginal prior of r and so the Gateaux derivative of the LRSE is always 0. Of course this is not true of the posterior mode which arises from the observed surprise and the principle of least surprise. Example 6. Suppose that x ~Bernoulli(0 1 / 2 ) , we take II to be the Beta(l,3/2) distribution and the support measure v is Lebesgue measure. Then the posterior distribution, when we observe x — 1, is given by the Beta(3/2,3/2) distribution and this has its mode at 1/2. Suppose we take Q to be a Beta(l, 1/2)

105

distribution and this implies that the posterior distribution is Beta(3/2,1/2) when we observe x = 1. Note that

el (1 e) 2+

e 2 (1 er 2

-' <» i*>=o - -» m m " - " "mm " - ' and this always has its mode at 1. This is an extreme example of course but it illustrates an important point. In particular the Gateaux derivative of the LRSE is 0 while, in this case the Gateaux derivative of the posterior mode is infinity. 4

Conclusions

In Evans x relative surprise inferences were advocated for Bayesian contexts where a loss function was not prescribed. One of the advantages of these inferences is that they are invariant under reparamerizations and this is not the case for more traditional approaches to deriving inferences in Bayesian inference problems. Further the relative surprise inferences are seen to be primarily data driven in that they are based on how the data changes beliefs from a priori to a posteriori rather than being based on the posterior alone. It might be argued that for a specific application a particular parameterization is paramount but reasons must be supplied for this and these do not seem available in many applications. Further it might be argued that if one has a strong belief in a particular prior then the relative surprise approach weakens the amount of input that the prior has in determining an inference. While this is true, the results of this paper point to the negative side of that argument, namely, the lack of robustness to the choice of the prior for some traditional inferences. The development here has excluded consideration of improper priors. Strictly speaking improper priors are excluded from the relative surprise formulation as they do not marginalize appropriately. Still we can consider many improper priors as limits of sequences of proper priors and consider the limiting surprise and relative surprise inferences as was done in Evans l . In such a case we could consider perturbations to this sequence and study the relative robustness properties of the limiting inferences. This is something we are currently investigating. This paper has only considered linear perturbations to the marginal prior. More generally we will study other types of perturbations. Further there is the question of the effect of perturbing the conditional prior distribution of the full parameter given the marginal parameter. It seems reasonable to separate out the effect of perturbations to the marginal and conditional. We will consider these questions in further research work.

106

Acknowledgments The authors thank the referees for some helpful comments. The authors were partially supported by the Natural Sciences and Engineering Research Council of Canada. References 1. M. Evans, Communications in Statistics - Theory and Methodology 26(5), 1125 (1997). 2. M. Evans, D. A. S. Eraser and G. Monette, Canadian Journal of Statistics 13, 137 (1985). 3. M. Evans and T. Zou, On the robustness of relative surprise inferences to the choice of prior, (Technical Report No. 0201, Dept. of Statistics, University of Toronto, Toronto, Canada, 2002). 4. P. Gustafson, Annals of Statistics 24, 174 (1996). 5. P. Gustafson, Journal of the American Statistical Association 9 1 , 774 (1996). 6. P. Gustafson, and L. Wasserman, Annals of Statistics 23, 2153 (1995). 7. R. E. Kass and A. E. Raferty, Journal of the American Statistical Association 90, 773 (1995). 8. W. Rudin, Real and Complex Analysis, Second Edition ( McGraw-Hill, New York, 1974).

U S I N G SURVIVAL ANALYSIS IN P R E T E R M BIRTH S T U D Y CHONG YAU FU The Institute of Public Health, National Yang Ming University, 155 Li-Long St., Sec. 2, Shih-Pai, Taipei, Taiwan, 112 E-mail: chong@ym. edu. tw SHIH HUA LIU Department of Humanities and Science, National Yulin University of Science and Technology, 123 University Road, Touliu, Yunlin, Taiwan, 640 In conventional preterm birth studies the outcome variable was a binary data, namely whether or not the gestational age was less than 37 weeks. An absolute yes or no classification was used. Based on the progressing sense of gestational age, this study investigated the risk of preterm birth among different age groups using survival analysis. Using the key concept of the rank ordering of time, KaplanMeier estimate and the Cox model were applied to preterm birth studies. The applied results show that the relative risk of preterm birth for the 12-17, 18-19 and 20-24 year old age groups, varied with gestational age, and was 1.6-8.69, 1.2-5.76 and 0.92-1.52 respectively when compared with 25-29 years old. For the older age groups, namely 30-34, 35-39 and 404- years old, the relative risks of preterm birth were 1.2, 1.54 and 1.74, respectively, constant in gestation age. Hence, applying survival analysis to preterm birth studies can provide more informative results than other approaches.

1

Introduction

Preterm birth is the leading cause of neonatal death and of morbidity in surviving infants ( Rush, Davey and Segall 10 ). The development of a complement system in preterm births was closely related to gestational age (Wolach et al. u ) . Hence, the duration of gestational age was important for preterm birth. Previous studies of preterm birth, namely Fu et al. 5 , Meis et al. 9 , Rush et al. 10 etc., always used a binary data approach, merely considering whether or not gestational age was less than 37 weeks. Naturally, the study results thus failed to reflect the relationship between risk of preterm birth and gestational age. Chiang 1 constructed an antenatal life table for fetal death and live birth in a prospective pregnancy study. The birth data was actually a history cohort data with the special feature of specific birthdays, the gestational age was a duration time of a fetus and the gestational age of less than 37 weeks was treated as an event. Hence, it should be straightforward to use survival analysis for a preterm birth study.

107

108 The censoring of data was a major issue in survival analysis. This study explores the performance of partial likelihood for the Cox model through censored data defined in this preterm birth study. The analytical results were expected to be more informative than those of other studies. 2

Statistical M e t h o d

Kaplan-Meier 7 estimate is an empirical method for estimating survival function. The Cox 2 model is commonly used for survival model incorporating covariates based on the partial likelihood function. Crowley and Hu 3 employ time-varying covariates in the Cox model. Both Kaplan-Meier method and Cox model are based on the concept of the rank ordering of time rather than on time itself. According to this concept, the units of time employed might be days, months, hours, gestational age, and so on. In this work, gestational age was equivalent to time, and birth at 37 weeks or less was an event. Births at 37 weeks or later were treated as censored data. The preterm birth data were then analyzed through survival analysis. Furthermore, graphs explored the original data and presented the modeled results. This study used the software, SAS 6.12 and S-plus (only for drawing plots). 3

Preterm Birth Study

Preterm birth is the cause of neonatal death, and of morbidity. Wolach et al. (1997) reported that the development of a complement system in preterm births is closely related to the gestational age . Hence, the conditions of preterm births varied with gestational age. And, the methodology of survival analysis is able to keep this information. Meis et al. 9 reported that the risk for preterm birth in different age groups displayed a U-shape, with younger and older mothers being high risks groups for preterm birth. In this study population, comparing with the 25-29 year age group, the relative risk for preterm birth in different age groups would be estimated. The data in this study came from the Computerized Medical Birth Registry (Lu et al. 8 ) , a data bank collected prospectively from ten Taiwanese hospitals from February 1993 to February 1995. The recruitment criteria included single birth, first parity, gestational age > 24 weeks, and birth weight < 500g. Totally, 17958 births met these criteria. Gestational age were estimated from the last menstrual period, and births with gestational age > 24 weeks and < 37 weeks were considered preterm

109 births and denned as event data. Those normal births, births with gestational age > 37 weeks, were treated as censored data. Age groups were classified as 12-16, 17-19, 20-24, 25-29(reference group), 30-34, 35-39 and 40+ years old. From a previous study (Fu et al. 5 ) , we realized that the risk factors for preterm births differed among younger and older age groups. Hence, the data set was divided into two parts, corresponding to younger and older age groups, both sharing a common reference group (25-29 years old). This approach increased the homogeneity each data set and did not introduce extra covariates into the model. However, the estimated coefficient values for the main risk factors (younger and older age) changed very slightly when additional covariates were brought into the models. 4

Results

Table 1 lists the number of preterm births and censored values for 24 — 37, 38 + weeks. Most of the births occurred in the 20 — 34 year old age group. Almost all censored data occurred at 38 + weeks. Figure 1 show the smoothed curves for the hazard of preterm birth over different age groups. These smoothed curves can be divided into three sections, for 24 — 28, 28 — 33, and 33 — 37, weeks in both plots of Fig.l. In the 24 — 28 week section, all curves display a constant hazard, except for those of the 12 — 17 and > 40 years age groups. Across all age groups, the hazard of preterm birth remained roughly constant at 28-33 weeks, but increased tremendously for gestational age > 33 weeks. Table l.The Frequency of Fails (preterm birth) and Censoring, (#failed/#censored), in Gestational Age 24 — 37 Weeks by age groups.

Weeks 24~ 24 25 26 27 28 29 30 31 32 33 34 35 36 37 36 +

12 - 17 yrs. 71 = 163 0/0 3/1 0/1 0/0 0/0 I/O 0/0 3/0 1/0 2/1 3/0 2/0 6/0 8/0 12/0 0/119

18 - 19 yrs. n = 384 0/0 0/0 5/0 0/0 2/0 1/0 2/0 1/0 4/0 4/0 4/0 13/0 7/0 8/0 31/0 0/302

20 - 24 yrs. n = 3, 545

25 - 29 yrs. n = 8,597

30 - 34 yrs. n = 4, 188

35 - 39 yrs. i% = 909

0/0 7/0 3/1 2/0 2/0 9/0 8/2 6/2 12/2 13/0 25/2 30/1 51/2 78/2 220/0 0/3069

0/0 6/4 7/2 7/0 12/0 12/1 10/0 16/4 32/4 34/4 53/1 54/1 85/2 213/0 586/3 0/7448

0/0 2/0 4/1 5/0 4/0 9/2 11/0 13/2 9/1 15/1 21/0 35/1 65/0 120/0 336/1 0/3529

0/0 1/0 1/0 1/0 2/0 1/0 3/0 4/0 6/0 8/0 5/0 10/0 14/0 40/0 83/1 0/729

TI

40+ yrs. = 172

0/0 1/0 0/0 1/0 0/0 0/0 1/0 0/0 0/0 2/0 1/0 1/0 5/0 5/0 21/1 0/133

Tables 2 and 3, and Fig. 2, summarize the results of Cox model. In table 2, model 2 provided significant time dependent results for younger age groups. Compared to the 25 — 29 year old age group, the relative risk of the younger 12 —17,18 —19 and 20 — 24 age groups varied with gestational age, and ranged

110 Table 2. Estimated Cox Models for Gestational Age in preterm births; Age Groups: 12-17, 18-19, 20-24, 25-29(reference group); Standard Errors in Parentheses

Covariates A l : 1 2 - 17 vs. 2 5 - 2 9

Model 1 0.768 (0.159)

Model 2 5.281 (1.498)

A2: 18 - 19 vs. 25 - 29

0.516 (0.116)

4.631 (1.164)

A3: 20 - 24 vs. 25 - 29

0.005 (0.055)

1.293 (0.711)

Al*Gestational Age

NA

-0.130 (0.044)

A2*Gestational Age

NA

-0.118 0.034

A3*Gestational Age

NA

-0.037 (0.020)

Total sample size: 12, 675; Events: 1,712; Censored: 10,963 (Censoring Rate = 86.5%)

Table 3. Estimated Cox Models for Gestational Age in preterm births; Age Groups: 25 — 29(reference group), 30 — 34,35 — 39,40+ ; Standard Errors in Parentheses

Covariates A5: 3 0 - 3 4 vs. 2 5 - 2 9

Model 3 0.179 (0.049)

Model 4 -0.119 (0.686)

A6: 35 - 39 vs. 25 - 29

0.441 (0.081)

1.018 (1.050)

A7: 40+ vs. 25 - 29

0.567 (0.165)

0.833 (2.184)

A5*Gestational Age

NA

0.008 (0.019)

A6*Gestational Age

NA

-0.016 0.030

A7*Gestational Age

NA

-0.008 (0.062)

Total sample size: 13, 849; Events: 1,991; Censored: 11, 858 (Censoring Rate = 85.6%)

111

Oil

0.11 12<=AGB=17 18<=AGB=19 20<=AGB=24 25<=AGB=29

ore

25<=ACB=29 30<=AGB=34 35<=AGB=3S 40<=ACE

DM

Off? -

0B7

•p

ra ^ 1155

016

ore

om

Ofll

-001

0D1

-T

24

1

1

oni

r-

26 28 30

32 34 36

38

— i

1

1

24 26 28 30

Gestational age (weeks)

1

1

1

32 34 36 38

Gestational age (weeks)

Figure 1: Smooth Hazard Functions for different age groups

12-17 y 18-19 y 20-24 y 25-29 ^(reference group) X-34y 35-38 y 40+ y

Ges&ional age (weeks)

Figure 2: Results from Cox models; age specific R.R. (reference group 25-29 yrs.) for preterm birth in gestational age 24-37 weeks.

112

from 1.6 — 8.67,1.2— 5.76 and 0.92 — 1.52, respectively. However, the relative risk for the older 30 — 34,35 — 39 and 40 + age groups did not vary significantly with gestational age (model 4 of table 3), and their relative risks were 1.2,1.54 and 1.74, respectively. Figure 2 displays the relative risk in younger and older age groups comparing with the 25 — 29 year old group. Comparing the empirical results (Fig. 1) with the modeled result (Fig. 2), the common impact of risk of preterm birth were the risk for younger age groups varying with gestational age while for older age groups this risk was constant. Meanwhile, the estimated relative risks were very significant for both models. 5

Discussion

The main methods employed in this application study were the Kaplan-Meier estimate and Cox models. Besides the key issue of censored data, the basic issues included time rank ordering and risk set. The appropriateness or otherwise of the application of the model can be judged based on these issues. This study is an original work applying survival analysis to investigate preterm birth. Gestational ages > 24 weeks and < 37 weeks were considered event data, and their ranked order was analyzed. Meanwhile, a gestational age of greater than 37 weeks was censored data, contributing to the risk set of denominators in parameter estimation, occurring in the survival estimator of Kaplan-Meier or the partial likelihood of the Cox-model. Consequently, the estimated parameters are based on the entire study population. Based on 17,958 births, the results of log-rank trend testing (Collett 4 ) indicated that ranked by age group, the risk for preterm birth was ordered as follows: (12 - 17) > (18 - 19) > (40+) > (35 - 39) > (20 - 24) > (30 - 34) > (25 — 29), with p — value < 0.0001. This testing was based on the sense of average risk in weeks 24 — 37. However, the relative risks of preterm birth varied with gestational age from Cox models for younger age groups. Hence, compared to the results of Cox models and the logrank test, Cox models not only dealt with the time dependent problem, but also estimated their relative risks. The risk factors for preterm births included maternal age, marital status, education, parity, prenatal care, and so on. Model building process revealed that maternal age was a very strong risk factor for preterm birth. Furthermore, with the very large sample size herein, whether or not other risk factors were included, the coefficient estimations of maternal age made very little difference. Consequently, this investigation only included the maternal age in Cox models.

113

However, the risk factors for preterm birth differed between the younger and older age groups. Hence, a sufficiently large sample size would offer the benefit of homogeneity in dealing with younger and older age groups separately. Finally, from two separate Cox models, the relative risk of preterm birth varied significantly with gestational age in the younger age group, but not in the older age group. Figure 1 does note this same situation. Reviewing previous studies, including Meis et al. 9 , Fraser et al. 6 , and Wessel et al. 12 preterm birth was defined as less than 37 weeks, thus classifying it as a binary response data. This work considered each gestational age using a survival analysis approach and thus incorporated a sense of progressive risk. Hence, these results were more informative than those that used a binary response. Acknowledgments I would like to thank Dr. Jen Her Lu for providing data from Computerized Medical Birth Registry and for useful discussion. References 1. C.L. Chiang , In The life table and its applications (Krieger, Florida, 1984). 2. D.R. Cox, Journal of the Royal Statistical Society, Ser. B 74, 187 (1972). 3. J. Crowley and M. Hu, Journal of American Statistical Association 78, 27 (1977). 4. D. Collett, In Modeling Survival Data in Medical Research, eds.: C. Chatfield and J.V. Zidek (Chapman and Hall ,London, 1994). 5. C. Y. Fu et al., Chinese J. of Public Health (Taipei) 18(3), 228 (1999). 6. A.M. Fraser, J.E. Brockert and R.H. Ward, The New England Journal of Medicine 332, 1113 (1995). 7. E.L. Kaplan and P. Meier, Journal of American Statistical Association 53, 457 (1958). 8. J.H. Lu et al., Medical Information 19(4), 323 (1994). 9. P.J. Meis et al., American Journal of Obstetrics and Gynecology 173, 590 (1995). 10. R.W. Rush , D.A. Davey and M.L. Segall, British Journal of Obstetrics and Gynecology 85, 806 (1978). 11. B. Wolach et al., Acta Padiatrica 86(5), 523 (1997). 12. H. Wessele£ al., Acta Obstetricia Gynecologica Scandinavica 75, 360 (1996).

A S Y M P T O T I C FORMS A N D B O U N D S FOR TAILS OF C O N V O L U T I O N S OF C O M P O U N D G E O M E T R I C D I S T R I B U T I O N S , W I T H APPLICATIONS JUN CAI Department of Statistics and Actuarial Sciences, University of Waterloo, Canada N2L 3G1 E-mail: [email protected] JOSE GARRIDO Department of Mathematics and Statistics, Concordia University, Quebec, Canada H^B 1R6 E-mail: garrido @vax2. concordia. ca

Ontario,

Montreal,

In this paper, we derive exponential and subexponential asymptotic forms for tails of convolutions of compound geometric distributions. General lower and upper bounds for the tails are given, which can be used in some cases to determine a closer tail approximation. Applications of these results are given to the ruin probability in the classical risk process perturbed by a diffusion; previous results are easily derived and a theorem of Veraverbeke 26 is generalized. In addition, twosided bounds for the ruin probability with large claim sizes are given, extending the bounds of Dickson 10 to the diffusion risk model.

1

Introduction

Let {Xi,i > 1} be a sequence of i.i.d. non-negative random variables with common distribution F and F(0) = 0. Further, let iV be a geometric random variable with Vr{N — n} = qpn, n = 0 , 1 , 2 , . . . and p = 1 - q, for 0 < q < 1, which is independent of {Xi,i > 1}. SN = J2i=i %i 1S sa*d to be compound geometric, where SN = 0 if N = 0. Its distribution function is denoted by H{x) = Pr{SV < x}, x > 0. Suppose that Y is another non-negative random variable with distribution G and (7(0) = 0, where Y,N and {Xi,i > 1} are independent. Then, the convolution H * G of the compound geometric distribution H and distribution G, i.e. the distribution of SN + Y, arises in many applied probability models, such as regenerative processes (Cohen 8 , Kalashnikov 18 and Keilson 19 ) risk theory (Dufresne and Gerber n , Sundt and Teugels 23 and Veraverbeke 26 ) and queueing theory (Asmussen 1 , van Hoorn 25 and Szekli 24 ). Many distributions of interest in these works can be expressed in the form H*G of the distribution of SN+Y. _ Denote by W(x) = H * G(x), consider W(x) = 1 - W(x), the tail or sur-

114

115

vival function of W(x). It is well-known (Renyi's theorem) that if F has a finite mean, then for any x > 0, H(x)/ /0°° H(y) dy ->• e~x as q -> 0 or equivalently £?(AT) —• oo. Furthermore, Keilson 19 (see also Kalashnikov 18 ) showed that if F has a finite second moment and G has a finite mean, then a similar limit theorem holds for W(x), namely for any x > 0, W(x)/ J0°° W(y) dy —> e~x as E(N) —> oo. However, in many applications, we are interested in the asymptotic behavior and bounds for W(x). The asymptotic behavior and bounds for the tail H(x) of the compound geometric distribution function H(x) is well-known. The purpose of this paper is to give a more complete description of the asymptotic behavior for W(x), obtain the asymptotic estimates for W(x) under various situations, derive bounds for W(x) in heavy tailed cases and consider the applications of these results in risk theory (for bounds see e.g. Cai and Garrido 4 ) . The paper is organized as follows: in Section 2, we derive an exponential asymptotic form for W(x) in terms of Lundberg's coefficient using the key renewal theorem. In Section 3, we consider subexponential asymptotic forms for W{x). Here, general lower and upper bounds for W{x) are given first, the bounds indicate the possible asymptotic form for W(x) and are also used to determine a closer approximation for W(x) in some cases. Section 4 discusses the asymptotic forms for W(x) in the intermediate case, when exponential moments exist but Lundberg's coefficient does not. In Section 5, as an application of the results for W(x), we consider the ruin probability in the classical process perturbed by a diffusion (see Rolski et al. 2 2 ) . The asymptotic estimates of the ruin probability derived by Dufresne and Gerber u , Gerber 16 and Veraverbeke 26 ) are easily obtained. A theorem of Veraverbeke 26 is also generalized. In addition, two-sided bounds for the ruin probability with large claims are given, thus extending the bound of Dickson 10 to diffusion risk models.

2

Light-tail asymptotics

In general, W(x) does not admit an exponential asymptotic form, for example, see Remark 3.5 of this paper. But if conditions similar to those in CramerLundberg's theorem hold, then there exists an exponential asymptotic form for W(x), which is stated in the following theorem. Throughout this paper, f(x) ~ g(x) means that f(x)/g(x) -> 1 as x -> oo and f(x) = o(g(x)) means that f(x)/g(x)-*0 as x—>oo while me^s) = /0°° esx dB{x) denotes the moment generating function of B on [0, oo).

116

Theorem 2.1. Let F be non-lattice and K a constant such that /•OO

eKx dF(x) = 1/p .

/

(1)

If mG(K) = /0°° e Kx dG(z) < oo then | ' i ) e - f a , if /? = / ° ° ie K » dF(a:) < oo , (2) P«P Jo = o(e-Kx), if ,9 = oo, (3) where, K; in (1) is called Lundberg's coefficient. Proof. See Willmot and Lin 28 , Corollary 9.3.2., pp. 175-6. • Remark 2.1. When Y = 0, G is degenerate at zero, and thus m o ^ ) = 1, W(x) = H(x) and Theorem 2.1 is reduced to Cramer-Lundberg's theorem, i.e. under the conditions of Theorem 2.1 about F H(x)~^e-Kx, if^
W

= o(e-Kx),

3

if/3 = oo.

(5)

Heavy-tail asymptotics

In this Section, we consider the case when F or G are heavy-tailed, in particular, subexponential distributions. First, the two following theorems give general lower and upper bounds for W(x), which indicate possible asymptotic forms used to determine, in some cases, a closer approximation for W(x). Theorem 3.1. For any x > 0,

ww >_ " F ( ; ' + f w .

(6)

pF(x) + q Proof. Since W{x) = PT{SN + Y > x} is the survival function of the random variable Sjy+Y, W(x) is decreasing. By the usual renewal argument, we have W(x) > qG(x)+pF(x)+PW{x)

f dF(y) ,

x >0,

Jo

= qG(x) +PF(x)+pW{x)F(x) this implies that (6) holds.

,

•

117

Remark 3.1. Take Y = 0 in Theorem 3.1, we get a lower bound for the tail H(x) of the compound geometric distribution function H(x), namely H(x)

>

f f ^ , for any* > 0 . (7) pF{x) +q This gives a derivation of a result known for the ruin probability in the classical risk process [e.g. see Theorem 3.1 of De Vylder and Goovaerts (1984)]. Theorem 3.2. If F has a finite mean E{X\) and G(x) has a decreasing density function, then for any x > 0 W(x)

< Pn*)+jG(x) pF(x)+q

+

6(x)

where S(x) = pG(x){x~1 E(Xi) - F(x)}-±0 as z->-oo. Proof. Since W(x) = H * G(x) = Pr{S;v + Y < x}, by conditioning on the value of Y, we get that for any x > 0, W(x) = G(x) + [X H(x - y) dG(y) Jo

(9)

[XH(x-y)G'(y)dy. (10) Jo But, we know that if f(x) and g(x) are integrable functions with different monotonicity on [a, b], a < b, then = G(x)+

pb

-i

f(x)g(x)dx Ja

rb

< —— / "

a

rb

f(x) dx

Ja

g(x)dx.

(11)

Ja

Thus, by (10), (11) and the fact that H(x — y) is increasing in y over [0, x], we get that for any x > 0, W{x) < G(x) + - T ff (a: - i/) dy [X G'(y) dy x Jo Jo - G(i) + ^ X

(12)

rI Hit), Hit) dt Jo /"OO

= G ( x ) + G W E(SN) - / (13) ^(0 X Jx Since 5jv is compound geometric, the distribution H of SV is a New Worse than Used (NWU) distribution [Lemma 2.1 of Brown (1990)], which implies that if is a NWUE distribution, i.e. />oo

L

H(t) dt > E(SN)H(x),

x>0.

(14)

118

Thus, by (13), (14) and (7), we get that for any x > 0 W{x) < G(x) + x-1G(x)E(SN)[l

- H(x)]


?3S«.

(15)

pF(x)+q But E(SN) = E(£%=i Xi) = E(N)E(Xi) = pE(X{)lq. Hence qE{SN) = pE(Xi). This, together with (15), implies that (8) holds. • R e m a r k 3.2. The requirement in Theorem 3.2 that G has a decreasing density function is not restrictive. In fact, this condition can be satisfied by many distributions such as the class of the equilibrium distribution function Fe(x) = J0 F(y)dy/ J0 F(y)dy with decreasing density function f(x) = F{x)/ Jg00 F{y) dy, which often arises in risk theory, reliability, queueing and renewal theory. In addition, the class of decreasing failure rate (DFR) distributions have decreasing density functions since f(x) = r(x) F(x), where r(x) > 0 is the failure rate function of F, which is decreasing in this case. Corollary 3 . 1 . Combining Theorems 3.1 and 3.2 we get that under the conditions and notation of Theorem 3.2, for any x > 0 PF(x)+qG(x) PF(x)

+q

_

<

pF(x)+jG(x)+S(x) pF(x)+q '

<

~

~

^

;

Since S(x) -> 0 as x -> oo, (16) indicates that PF{x)

+ gG(x) pF(x)+q

may be an asymptotic form for W(x) as x -+ oo. Indeed, we show below that under various situations, this is precisely the asymptotic form of W{x). Corollary 3.2. Under Theorem 3.2, if lim xG(x) = oo then x-+oo

pF{x)+q Proof. By (16), we get for any x > 0,

! <

IT(X)

W(x) v J

pF(x)

+

J

+ qG{x)

< 1+ ~

PGMiEm-xFjx)} PxF{x)

+ qxG{x)

v

119

Since F has a finite mean, x F(x) -t 0 as x -> oo. Thus, (18) and x G(x) -> oo as x —> co imply that W(x)

~ PFW + qG(x) pF(x)+q D

R e m a r k 3.3. There exist classes of distributions that satisfy the conditions of Corollary 3.2, for example the Pareto distribution with density function q(x) — -

aca —— ,

x >0,

where 0 < a < 1 , c > 0,

or more generally, the Burr distribution with distribution function F(x) = 1 - ( T

)

,

x >0,

where A > 0 , 0 < a < l , 0 < r < l .

But, it should be pointed out that the condition xG(x) -^ oo as a:-»oo is restrictive, since it implies that G has no finite mean. On the other hand, under the conditions of Corollary 3.2, pF(x)+gG(x) PF(x)+q gives a closer approximation to W{x) than G(x) does, as seen by Theorem 3.1 and the fact that for any x > 0,

pnx)+iG(x) PF(x)

+q

>

~

We know that if h is an asymptotic form of an unknown function T, i.e. T(x) ~ h(x), then any function g satisfying g(x) ~ h(x) is also an asymptotic form of T. Hence, it is interesting to find a closer asymptotic form among known approximations. As shown above, a closer approximation can be derived by combining bounds and asymptotic forms. Furthermore, we notice that the only relation between F and G required in Corollary 3.2 is that G(x) dominate asymptotically F{x). If other relations between them are assumed and subexponentiality (as defined below) is further imposed on F or G, we can derive additional results for W(x) as follows. Definition 3 . 1 . A distribution B on [0, oo) is said to be subexponential, denoted by B 6 S, \iB~$){x) ~ 2~B{x). Subexponential distributions are heavy tailed; typical examples are the Pareto or Lognormal distributions. The following Lemma combines Proposition 1 of Embrechts et al. 14 and Theorem 2 of Chistyakov 5 .

120

Lemma 3.1. Suppose that F\ and F 2 are two distributions on [0, 00). (i) If F 2 G S and Fx{x) = o{F2(x)), then Fi*F 2 G S and Fi * F2(x) ~ F 2 (x). (ii) If Fi * F2 G 5 and F^x) Fi*F 2 (ar).

= o(Fi *F 2 (x)) then F 2 G 5 and F2(x)

(iii) If F 2 G 5 , then for any e > 0, lim eSxft(x)

~

= 00, i.e. e~£x = o(Fj(a;)).

Theorem 3.3. Suppose that G(a;) = o(F(x)). The three following assertions are equivalent: (i) F 6 S, (ii) W (iii) W(x)

eS, ~ { p f ( a ; ) + g G ( x ) } / { p F ( x ) + 9} ~

pF(x)/q.

Proof. First, it is clear that G{x) = o(F(x)) implies pF(x)+qG(x) pF(x)+q

^

By Corollary 3.2 of Embrechts et al. conditions are equivalent:

14

pF(x) q

, we know that the following three

(a) FeS, (b) H G S, (c) i/fa) ~ pF{x)/q . Hence, if F G S, by (b), (c) and G{x) = o(F(a;)), we get that H G S and

£W=EwxM40 H(x)

F{x)

as

^00j

H{x)

i.e. G(:r) = o(H(x)), hence (c) and (i) of Lemma 3.1 imply that W = G*H G S and W(x) ~ i? (x) ~ pF(x)/q. Conversely, if W = G * H G S, by Theorem 3.1 and G(x) = o(F(a;)) : G_(x) < { ^ ( x ) + g g f r ) - H ' ( i ) - pF(i)+?G(i) <

=

{pF(x) + j } G ( z ) / F » p + 9 G(a:)/F(a;)

_^

Q

this is to say that G(x) = o(W(x)), thus, (ii) of Lemma 3.1 and (c) imply that H G 5 and W(x) ~ # (a:) ~ pF(x)/q. So, we have shown that F G <S 44> W e S. In addition, the above proof also showed that F G S =$• W(x) ~ pF(x)/q. Thus, in order to complete the proof of Theorem 3.3, we still need to prove that W(x)

~ -F(x)

=>• FeS.

(20)

121

By definition, W{x) = Pr{5jv + Y <x} = £ ~ = 0 qPnG * F^(x),

and hence

^qpnG*F(n){x)

W{x) = n=0

which implies that G*FW(x)

= —; w{x)-Y^qpnG*F^{x) qp' n^2

(21)

Since Y is a non-negative random variable, for any integer k > 1, G * FW(x) = Pv{Y + X1 + --- +

Xk>x} F(k)(x)

> Pr{X 2 + • • • + Xk > x} = >Pr{max(X1;...

,Xk)>x} k-i

k

F(x)Y,[F(x)]n

= l-[F(x)] =

(22)

n=0

(22) implies that hm mf

_

x - > oo

F(x)

w N

> lim inf

_ / /

z ->• oo

> k.

(23)

F(x)

Clearly, (22) and (23) are also true for k = 0, thus by (21) and (22) FW(x) F(x)

G*F(V(x) ~ F\x)

1 - qp^

Thus by (24), (23) and W(x) ~ pF(x)/q,

W(x)

^

qp

W)-^2

F(x)

(24)

we get that

FW(x) 1 hmsup - = — — < —r-~2^
y

n^2

<


q

*—'

qp'

+ 2
= 2 , (25)

n=0

hence, (23) and (25) imply that Iim^ -> oo i ^ 2 ' (x)f F(x) = 2, i.e. F e S. a It is interesting to note that (20) holds and is independent of the condition that G(x) — o(F(x)). In addition, if G is an exponential distribution, using Lemma 3.1(iii) and following the proof of Theorem 3.3, we get directly the following corollary.

122

Corollary 3.3. If G is an exponential distribution, then Theorem 3.3 holds. Theorem 3.4. Suppose that F(x) — o(G(x)) and F E S. The two following assertions are equivalent: (i) GeS,

(ii) W G S,

and either one of them implies that: (iii) W{x)

~

{PF{x)+qG{x)}/{pF{x)+q}

~

G(x).

Proof. By (c) in the proof of Theorem 3.3, we know that H(x) = ^ -

H(x)

F(x)

= =7-7- x = ^

-)• 0

as

x -> oo ,

G{x) F(x) G(x) i.e. H(x) = o(G(x)), hence if G G S, by Lemma 3.1(i), we get that W = # * G G S and W(x) ~ G(z). But, F(z) = o(G(x)) implies that pF(z) + g G ( i ) m pF(x)+q Conversely, if W = H * G G S, by Theorem 3.1 and the proof of Theorem 3.3(c), we get that < £ ( * ) < {pF(x) + g f f l z ) - W(a;) ~ p f ( x ) + ? G ( i )

=

{pF(x) + q)E{x)IG{x) g F(x)/G(x)+
_^

&g

i.e. #(:r) = o(W(a;)). Then Lemma 3.1(h) =>• G G 5 and W(x) ~ G(z). Remark 3.4. We know that pF(a;) + oG(a;) pF(x) „ , , , 9^, . „ r _ F * V ; > £—i-^ if and only if g2 G(x) > \pF(x)}2. pF{x)+q q

•

Thus, in view of Theorem 3.3(iii),_if F G S^x) = o(F(x)) and q2G(x) > 2 3 2 \pF{x)} as a;->oo, (for example, G(x) = [F(x)] / ), then by Theorem 3.1, we know that {pF(x) + qG(x)}/{pF(x) + q) is a closer approximation for W{x) than pF{x)/q is. By an argument similar to that in Remark 3.3, we also know that under the conditions of Theorem 3.4 and if G G <S, then {pF{x) +qG(x)}/{pF(x)+q} is a closer approximation to W(x) than G{x). Theorem 3.5. If G(x) ~ T{x) and F G S, then w

W (x) ~

PF(x)

+ qG(x) =—— pF{x)+q

F(x) ~

G(x) ~

q

. q

123

Proof. F E S implies that H e S and H(x) ~ pF(x)/q. G(x) H(x)

G(x) F(x)

Fix) H(x)

Thus, by Theorem 1 of Cline Kliippelberg 17 ) we know that

6

q p

(see also Theorem 4.3 of Goldie and

q H(x) ~ (l + l)H(x) = p p On the other hand, F(x) ~ G(x) implies that W(x)=G*H(x)

PF(x)+qG{x)

F{x) q

pF(x)+q

Hence,

F(x)

G(x) q

so Theorem 3.5 holds. • Remark 3.5. The results of this Section also show that the lower bound in Theorem 3.1 is asymptotically exact for W(x) as x -» oo, in the subexponential case. Bounds for W(x) were considered in Kalashnikov 18 and Willmot and Lin 2 7 ) . The bounds for W(x) in (5.1) and (5.2) of Kalashnikov 18 are asymptotically exact as E{N) -> oo. The bounds of Willmot and Lin 27 are applicable to the tail of convolutions of more general compound distributions; these are based on a generalized Lundberg's coefficient and NWU distributions (see Willmot and Lin 28 for more details). Also, we point out that W{x) cannot admit exponential asymptotic forms and exponential upper bounds if F or G is subexponential, i.e. no constant c > 0 and e > 0 such that W(x) ~ ce~£x or W(x) < ce~£x, for all x > 0. For example, if F is subexponential, then we know that H € S and e£x H(x) —> oo as x -> oo, hence lim e£x W(x) = oo, since W(x) = H * G(x) > H(x). x—>oo

4

Medium-tail asymptotics

We consider here the intermediate case, i.e. when mF(s) < oo for some s > 0, but Lundberg's coefficient does not exist, as Tnp(s) = 1/p can not be satisfied but mpis) < 1/p holds (e.g. in the inverse Gaussian case). First, recall the definition of the S(a) class and its properties. Definition 4.1. A distribution B on [0, oo) is said to belong the S(a) class for a > 0, denoted by B S S(a), if (i) \imx^00BW(x)/B(x) (ii) limx _> oo B{x - y)/^(x)

=

2mB{a)
124

Clearly, B € 5(0) <£• B G S. A class of distributions in S(a) is the class of generalized inverse Gaussian distribution N~1(a,b,c) with a < 0, b > 0 and c > 0, since N~1(a,b,c) G <S(c/2) (see Embrechts 12 for details). The following proposition recalls some properties of <S(a), which will be used here. Its proof is in Lemma 2.4 and Theorem 2.7 of Embrechts and Goldie 13 and page 268 of Kliippelberg 21 (for details see Rolski et al. 2 2 ) . Proposition 4.1. Suppose that B G £(7). (i) For any e > 0, e 7 + e B(x) —> 00 as a; —• 00, (ii) If L is a distribution on [0, 00) and limx —> 00 L{x)/ B{x) = c, where 0 < c < 00, then L € 5(7), (iii) If 7 > 0, B has a finite mean m, and B\ is the ladder height distribution of B [i.e. B,{x) = i | 0 X B(j/)dy], then B x G «S(7) and ^ ( z ) ~ g £ } . Theorem 4 . 1 . Suppose that F G 5(7) for some 7 > 0 and that 771^(7) = | 0 °° e<xdF{x) < 1/p. If G(x)/F{x)-+a as z - ^ o o , then W G 5(7) and F(») ~ g W 7 ) + "[l-pmF(7)]} [l-pmF(7)J2 Proof. Since 0 < prnpij) < 1, there exists some e > 0 such that 0 < p[mF('y) +e] < 1, thus £^L 0
n 1

where c = Y^Li nqp [mF('y)] -

(27) 2

= pq[l - puip^)]' .

G{x) G(x) F(x) a —, ' = =i—^ x _ -> — as #(x) F(a:) H(i) c By Theorem 1 of Cline 6 , we get that W{x) hm = 7 ~ f = 1 -> 00 f f ( x )

,. H*G(x) hm x->oo

;

Thus, x -> 00 .

. , a = mG 7 + - * " g 7 •

H(x)

C

But, m^(7) = £ ( e ^ ) =

^ ^

n

E

afZ7=iXi

n=0

= X > " [ m F ( 7 ) ] " = q[l - p m F ( 7 ) ] - 1 n=0

(28)

125

Hence, by (27) and (28), we get that Q<

W(x) ~ [TTIGCT) H—wjf (7)] # ( z ) ~ [email) + amH(7)] 2

F(a;)

-1

= {p?[l-pmF(7)]- mG(7) + a 9 [ l - P " ^ F ( 7 ) ] } ^ ( Z ) . i.e. (26) holds. W £ S(j) follows from (26) and (ii) of Proposition 4.1. 5

•

Ruin probabilities in a diffusion risk model

To illustrate a possible application of the above results, consider the classical risk process perturbed by a Wiener process, i.e. the surplus R(t) at time t is R(t) = x + ct- S{t) + Z(t) , where x > 0 is the initial risk reserve, c > 0 is the premium rate, S(t) is the compound Poisson process representing the total claims at t [with rate 1/d > 0, and the independent individual claim sizes with common distribution function B and B(0) = 0], and {Z(t)} is a Wiener process, independent of {5(t)}, with infinitesimal drift 0 and infinitesimal variance 2D > 0. Assume c > A/d, where A = /0°° B{x) dx is the expected claim size, then the relative security loading q = 1 — \/(cd) is such that 0 < q < 1. Let ip(x) denote the probability of ultimate ruin, starting with initial reserve x : V>(a:) = Pr{infiJ(i) < 0} . Denote tp(x) = 1 — i(>(x). Dufresne and Gerber have shown that for any x > 0,

n

(see also Veraverbeke

26

)

00

tp(x) = Y, qpnF{n) * G(x) =H* G(x) , n=0

or, equivalently, 00

i(j(x) = J2 1Pn F(n) * G(x) = IlTG(x)

,

(29)

n=0

where H{x) = £ £ L p q p n F ^ ( x ) , G{x) = 1 - e~cx/D, F(x) = G * B1 (x), Bx [x) = \ $* B(y)dy, for x > 0, and p = A/(od), q=l-p=j=ft. Suppose that ^ and r) are independent random variables with distributions G and B\, respect., then £ + r\ has distribution F' = G*B\ and for any s £ l , />oo

mF(s) = / Jo

esxdF(x)

= £ [ e ^ + " > ] = E(esi)E(es»)

= mG(s)mBl

(s). (30)

126

Thus, if there exists a constant R such that mF(R) = mG(R)mB1{R)

(31)

=

then (31) implies that R < c/D , r

mG(R) = E(e"t) = -J

rc

Rx

e

— ^-

i

e o dx

c-RD

(32)

< oo

and xe

e

D

D mG{R) . c-RD

cD (c - RD)2

dx =

(33)

By (31) and (32), we get that E(eR*) = mBl(R)

=

1 pmG{R)

c-RD pc

Hence />00

/? = / Jo = 5.1

xeRxdF{x) mG(R)

pc

= E[{£ + V)em+rt)

A J0

= E{^)E{eRr])

+ E(e fi «)S(r/e fi ")

xeHxB(x)dx

(34)

Asymptotics for the diffusion risk model

We know that if G and B\ have density functions, so does F = G * B\. This implies that F is non-lattice. Thus by Theorem 2.1, we get the following exponential formula for the ruin probability tp(x). Corollary 5.1. Suppose that mp{R) = 1/p. If J0°° xeRx B(x) dx < oo then

«*>-=$?'-*- and if JQ xeRx B{x)dx

*E+*L r c

cd J0

Rx xe

B{x)dx

-l

-Rx

= oo, then ip{x) = o{e~Rx) .

(35)

This Corollary includes Theorem 4.1 of Gerber 16 , the results in Section 7 of Dufresne and Gerber u and in Section 3 of Veraverbeke 26 .

127

Since G{x) = 1 - e~cxlD is an exponential distribution function, by (i) and (ii) of Lemma 3.1, we know that B\ 6 S <S> F = G*Bi £ S. Furthermore, by Corollary 3.3 and Lemma 3.1, we know that Bi e S =S> 1>(x) ~ - F(x) = —^— T(x) q cd — X cd — A

cd-

In addition, using the fact F{x) = G * Bx(x) > Bi(x) and following the proof of (20), it is clear that %l>{x) ~ ? 57(3;) = _ ^ _ (J

B~1[X)

=> Bi e <S ,

CO — A

thus, Corollary 3.3 implies the following theorem. Theorem 5.1. For the classical risk process perturbed by a diffusion, the following conditions are equivalent: (i) Bx £ S,

(ii) v e S,

(iii) 4(x) ~ ^

/x°° B(y) dy.

Remark 5.1. Theorem 5.1 generalizes Theorem 1 of Veraverbeke 26 , where it is shown that (i) and (ii) are equivalent and that either one of them implies (iii). Conditions enabling B\ £ S can be found, expressed in terms of B, in Embrechts and Omey 15 and Kliippelberg 20 . Finally, for the intermediate case, suppose that for some 7 > 0, BE S(J) and 771^(7) = W G ( 7 ) J « B 1 ( 7 ) < 1/p, then by Proposition 4.1, we get that B\ £ S("f) and B\(x) ~ B(x)/(/y\), in addition, (1) rnG{l) = (c/D) J0°° e?xercxlD

dx = c/{c - D7) < 00,

(2) mBl (7) = Jo00 e 7 * dBt (x) = (1/A) J0°° e^ B~[(x) dx < 00 and 7 < c/D, hence, there exists some e > 0 such that 7 + e < c/£>, thus m g ( 7 + e) = c / [ c ~ (7 + e)JC] < °°) a n d by Proposition 4.1, we get that G(a;) e-y+£G(x) . = z=~ jBi(a:)

e-y+£Bi(x)

> 0 as a; -> 00 .

Thus, by Theorem 1 of Cline 6 and (ii) of Proposition 4.1, we get F(x) G*B1(x) = — = —= -> man) Bx{x) B1{x)

as

z-^oo

128

and F € S ( 7 ) and F(x)

~

mG^)B~[(x)

~

T

^^-B(x)

Thus, G(x) = o(F(a;)) and Theorem 4.1 imply that
*(*) ~ n W m G ( / \ 1 2? F ^

(36)

2

[l-pmF(j)] pg[m G (7)] 2 A7[l-pmF(7)]

= cdj — ( 1 -cd - ) l -

2

elyB{y)dy

-A c cd J0 thus, (38) yields Theorem 2 of Veraverbeke 26 . 5.2

B(x) ,

(38)

Ruin probability bounds for the diffusion risk model with heavy tails

Under condition (31), the exponential bounds for the ruin probability ip(x) have been derived by Dufresne and Gerber n . Here, we use a generalized condition of Dickson 10 to derive bounds for if)(x) with heavy claim size tails. General upper and lower bounds for 4>(x) follow directly by Theorems 3.1 and 3.2 since G is exponential and has a decreasing density. Given t > 0, suppose that Rt satisfies mG(Rt)

f eR'ydB1(y) Jo

= -, P

(39)

or equivalently,

I

* e*<*dBM = —— • (40) /o cp L e m m a 5.1. For any claim size distribution B with B(0) = 0, there exists a unique solution Rt € (0, c/D) to equation (39). Proof. Let h(x) = mG{x) /0* exydBi(y) - ±. Since H(0) = Bi{t)-±<0 and lim moix) = lim xtc/D

xfc/D

— = 00 , C — xD

this implies that limx1-c/£) h(x) = 00. Thus, the existence of the unique root of h follows from the fact that h is continuous and strictly increasing. • Since B\ is continuous in here, condition (39) is clearly equivalent to mam

J

^ydBt(y) = mry

(41)

129

where

mix) =

{BMmt)so^x
(42)

Let ^ W = ^ftPl»FfW*G(x)

(43)

n=0

or equivalently, %l)t {x) = l - tpt (i) = ^2qtPt

^ ( n ) * G{x) ,

(44)

n=l

where p t — pB\{t), qt = 1 - Pt and Ft — Bt* G. That is to say, Vt(^) i s the ruin probability in the diffusion risk model with corresponding parameters Pt, qt, Ft and G. By induction we get that for any 0 < x < t, B(tn)(x) = B[n)(x)/[Bi(t)]n, hence for any 0 < x < t,

J2^ptBtn)*Gin)*G(x)

n^) = n=0

oo

= - J2 QPnF(n) * G{x) = -
Q

Thus, for any 0 < x < t, 1>t{x) = 1 - ft{x) = 1 - - ipd{x) = 1 - ^ [1 - tl>d(x)] . This implies the following property. Lemma 5.2. For any 0 < x < t, ^

'

q + pBrit)

{

q + pB^t)

'

Using Lemma 5.2 above, we can prove the following result. Theorem 5.2. Suppose Rt satisfies (39), then for any 0 < x < t,

p*JfL- < ^

q+pB^t) ~ In particular, for any x > 0, P 1(X)

*

-

q + pBi{x)

< JML.

~ q+pBiit)

< iK«) <

P 1 X)

*A

q+pBi{x)

+

+

J!£±L..

(46)

q+pBi(t)

qe X

'-

q+pB^x)

.

(47)

130

Proof. The lower bound in (46) follows from ipt(x) > 0 and (45). On the other hand, if a constant R satisfies 1

f°°

mG(R) then Dufresne and Gerber

16

/ Jo

eR"dB^y)

= - , P

(48)

show that for any x > 0, ip(x) < e-Rx

.

(49)

Thus, apply (49) to ipt{x) with condition (41), to get for any x > 0, tPt(x) < e~R'x . This, together with (45), implies that the upper bound in (46) holds. Taking x = t in (46), gives (47). • Remark 5.2. (a) Since pB^x) q+pBx{x)

pB^x) q

by Theorem 5.1, we know that the lower bound in (47) is asymptotically exact, for large x, if £?i is a subexponential distribution. (b) If D = 0, the diffusion risk model is reduced to the compound Poisson risk model, thus the bound of Dickson 10 is derived as a special case of Theorem 5.2 and is improved upon. Other applications of the results in sections 2-4 have been investigated, notably to M/G/k queues (see for example Asmussen 2 ). Acknowledgments We are grateful to the anonymous referees for their constructive comments. We would also like to acknowledge the funding obtained for this research by the Natural Sciences and Engineering Council of Canada (NSERC) operating grant OGP0036860. References 1. S. Asmussen, Applied Probability and Queues (John Wiley & Sons, New York, 1987). 2. S. Asmussen, Ruin Probabilities (World Scientific, Singapore, 2000). 3. M. Brown, An. Prob. 18, 1388 (1990). 4. J. Cai and J. Garrido, J. App. Prob. 36, 1058 (1999). 5. V.P. Chistyakov, Th. Prob. App. 9, 640 (1964).

131

6. D.B.H. Cline, Prob. Th. Rel. Fields 72, 529 (1986). 7. D.B.H. Cline, J. Aus. Math. Soc. A43, 347 (1987). 8. J.W. Cohen, On Regenerative Processes in Queueing Theory (SpringerVerlag, Berlin, 1976). 9. F. De Vylder and M.J. Goovaerts, Ins.: Math. Econ. 3, 121 (1984). 10. D.C.M. Dickson, Scand. Act. J. , 131 (1994). 11. F. Dufresne and H. Gerber, Ins.: Math. Econ.10 51 1991. 12. P. Embrechts, J. App. Prob. 20, 537 (1983). 13. P. Embrechts and C M . Goldie, Stoc. Proc. App. 13, 263 (1982). 14. P. Embrechts, C M . Goldie and N. Veraverbeke, Z. Wahr. Verw. Geb. 49, 335 (1979). 15. P. Embrechts and E. Omey, J. App. Prob. 2 1 , 80 (1984). 16. H. Gerber, Skand. Akt. 2051970. 17. C M . Goldie and C Kluppelberg, in A Practical Guide to Heavy Tails: Statistical Techniques and Applications, ed. R. Adler, R. Feldman and M.S. Taqqu (Birkhauser, Boston, 1998). 18. V. Kalashnikov, Topics on Regenerative Processes, CRC Press, Boca Raton, Florida, 1994). 19. J. Keilson, Ann. Math. Stat. 37, 886 (1966). 20. C Kluppelberg, J. App. Prob. 25, 132 (1988). 21. C. Kliippelberg, Prob. Th. Rel. Fields 82, 259 (1989). 22. T.Rolski, H.Schmidli, V. Schmidt and J. Teugels, Stochastic Processes for Insurance and Finance (John Wiley & Sons, New York, 1999). 23. B. Sundt and J.L. Teugels, Ins.: Math. Econ. 16, 7 (1995). 24. R. Szekli, Stochastic Ordering and Dependence in Applied Probability, Lect. Notes Statist. 97 (Springer-Verlag, New York 1995). 25. M.H. van Hoorn, Algorithms and approximations for queueing systems, CWI Tract# 8 (CWI, Amsterdam, 1984). 26. N. Veraverbeke, Ins.: Math. Econ., 13, 57 (1993). 27. G.E. Willmot and X. Lin, Ins.: Math. Econ., 18, 29 (1996). 28. G.E. Willmot and X. Lin, Lundberg Approximations for Compound Distributions with Insurance Applications, Lee. Notes Stat. 156 (SpringerVerlag, New York, 2001).

I M P R O V E D FINITE-SAMPLE I N F E R E N C E IN O V E R I D E N T I F I E D MODELS W I T H W E A K I N S T R U M E N T S NIKOLAY GOSPODINOV Department of Economics, Concordia University, 1455 de Maisonneuve Blvd. West Montreal, Quebec, H3G 1M8 Canada E-mail: gospodin@vax2. concordia. ca This paper investigates the finite-sample properties of the class of generalized empirical likelihood estimators in possibly overidentified models with weakly identified parameters. These nonparametric likelihood estimators satisfy exactly the moment conditions and automatically remove the bias that arises from a lack of centering of the moment conditions. The inference procedure suggested in the paper does not involve any explicit estimation of the variance-covariance matrix. The confidence sets for the parameters of interest are constructed by inverting the x 2 acceptance region of the criterion test.

1

Introduction

Moment condition models arise naturally from dynamic economic theory with optimizing agents. Since the seminal paper by Hansen 13 , the generalized method of moments (GMM) has become the predominant framework for estimating the structural parameters of these models. Under some general regularity conditions, the GMM estimator is consistent, asymptotically normal and efficient for the given set of moment conditions. Unfortunately, it has been found that the small-sample properties of the conventional GMM estimators (in particular, the two-step GMM) are rather poor. In this paper, we investigate the properties of the class of generalized empirical likelihood estimators of moment condition models. Members of this class are the empirical likelihood-based GMM of Qin and Lawless 27 , Imbens 15 and Imbens, Spady and Jonhson 16 , and the maximum entropy-based GMM of Kitamura and Stutzer 17 and Imbens, Spady and Johnson 16 . This class also includes the continuously-updated GMM as a special case (Imbens, Spady and Johnson 16 ; Newey and Smith 22 ) which explains the superior small-sample performance of this estimator over the traditional two-step GMM found in Hansen, Heaton and Yaron 14 . These nonparametric likelihood estimators minimize the distance between the empirical distribution function and a distribution function that exactly satisfies the moment conditions. One of the most attractive properties of nonparametric likelihood estimators is that they tend to remove some important sources of bias that give rise to poor finite-sample properties of the GMM estimator and GMM-based

132

133

test statistics. The first source of bias arises from the fact that the first-order conditions of the standard two-step GMM estimator (Hansen 1 3 ) , evaluated at the true values of the parameters, are non-zero. This bias is exacerbated if the number of instruments increases (Kocherlakota 1 8 ). Altonji and Segal 1 and Angrist, Imbens and Krueger 3 proposed some ad hoc methods for reducing the magnitude of the bias. Donald and Newey 8 showed that, for the continuously-updated GMM of Hansen, Heaton and Yaron 14 , the first-order conditions are exactly satisfied at the true values of the parameters and this source of bias is automatically removed. In fact, for all members of the class of the generalized empirical estimators, the moment conditions are exactly centered at zero by construction. Second, the estimation of the weighting matrix can be another important source of bias due to the non-zero finite sample correlation between the elements of the variance-covariance matrix of the parameters and the errors (Altonji and Segal 1 ) . This source of bias is present for both the two-step and continuously-updated GMM estimators but disappears for the empirical likelihood (EL) estimator of Qin and Lawless 2T . Newey and Smith 22 showed that the bias of the empirical likelihood estimator of Owen 24 and Qin and Lawless 27 is the same as the bias for the infeasible optimal GMM where the optimal linear combination coefficients do not have to be estimated. Third, the small-sample properties of the GMM estimators and test statistics can be seriously affected by the choice of instruments that are only weakly correlated with the endogenous variables (Stock and Wright 2 9 ) . In this case, the finite sample distributions of the GMM estimators and the test statistics may depart substantially from their asymptotic distributions. Stock and Wright 29 proposed an alternative reparameterization of the moment conditions and obtained asymptotic representations with improved finite sample properties. In their framework, however, the weakly identified parameters are not consistently estimable. Fortunately, one could still conduct asymptotically valid inference by inverting criterion-based tests since their limiting X 2 -distribution at the true values of the parameters is preserved. In this paper, we show that the nonparametric likelihood estimators are robust in the presence of weakly identified parameters. Most importantly, the criterion-based inference procedure does not involve any explicit estimation of variance-covariance matrices. Unlike the Wald test, the confidence sets constructed by inverting the criterion test, also satisfy the requirement of infinite expected volume in the completely unidentified model (Dufour 9 ) . Finally, the class of generalized empirical likelihood estimators is transformation invariant and the obtained confidence sets are transformation respecting. The rest of the paper is structured as follows. Section 2 discusses two ap-

134

proaches to estimating moment condition models that give rise to the GMM and the nonparametric likelihood estimators. The asymptotic validity of the confidence interval construction by criterion test inversion is shown in Section 3. The Monte Carlo experiment in Section 4 studies the finite-sample properties of the different estimators and their corresponding confidence intervals in a linear instrumental variable model with weakly identified parameters. In Section 5, nonparametric likelihood estimators are applied to estimating the return to education. Section 6 summarizes the conclusions. 2

General Approach to Estimating Moment Condition Models

Let E[g(x,9)\F] = J g(x,6)dF = 0 be an m x 1 vector of population moment conditions implied by economic theory, where (xi, xi,...) are independent random vectors in R p with unknown continuous distribution function F, 8 is a k x 1 vector of unknown parameters from 0 and g(.) is a given function {g(x,0) : R p x R l - » R m } with m > k. Suppose that we restrict the family of possible distribution functions to the space of multinomial distributions with finite support on the observed data, denoted by $. Also, let Fn denote the empirical measure of the sample {xi}"=1 from F that places probability mass n _ 1 on each data point and P„ be another probability measure that assigns multinomial weights Pi,P2, ••••,Pn to each of the observations. Below, we consider two versions of the analogy principle discussed in Manski 2 0 . The first version selects an estimator that minimizes the distance of the moment conditions from zero (GMM estimators). The second version selects an estimator that minimizes the distance between the empirical measure and a measure Pn that satisfies exactly the moment conditions (nonparametric likelihood estimators). 2.1

GMM Estimators

The conventional GMM estimator minimizes the distance of the sample counterparts of these moment conditions from zero using the quadratic form Qn(0) = 9n(6)'Wn(6)gn(6),

(1)

where gn(0) = E[g(x,6)\Fn) = f g(x,9)dFn and Wn(9) is a positive definite weighting matrix. Then, 9 = argmme<E@Qn(6). The properties of the GMM estimator depend crucially on the choice of the weighting matrix. The optimal GMM estimator sets Wn{9) = [£ £ t " = i & ( % # ) ' ] ~*, where gi{9) = g{Xi, 6). If a preliminary consistent (but not necessary efficient) estimator 6 of 9 is

135

used in the estimation of Wn, we have the two-step GMM estimator Oistep = argmin ( ? € 0

• (2) «=1

J

L

i=l

.

.

«=1

If we substitute 82step in the weighting matrix and repeat this until convergence of both 9 and Wn, the obtained estimator is the iterated GMM estimator (Hansen, Heaton and Yaron 14 ) described by the equations -1

2_ V^ / d9i(&igmm) 56"

n

i=l

/ j 9i \yigmm)9i

("igmm )

9n\Pigmm)

— Ufc.

(3) Finally, the continuously-updated GMM estimator proposed by Hansen, Heaton and Yaron 14 does not require a preliminary estimate of 6 and directly minimizes the criterion function 6CU = argmin0 € O

• j=i

J L i=i

J

(4)

L «=i

The estimator is the solution to a (typically nonlinear) system of k first-order conditions I ^ j g H

I W J ^ J

5 n

( e

c

0 - 5 „ ^ c „ ) ' W „ ( ^ n ) ^ ^ W n ( ^ ) 5 n ( ^ )

= 0

or
J r

^

-v

L I-i

t=l ^

where A = - [ £ " = i 5i(^cu)3i(^cu)'J 9n(0cu)Although these estimators are asymptotically equivalent, their finite sample properties may differ (see for example Hansen, Heaton and Yaron 1 4 ). 2.2

Nonparametric Likelihood Estimators

A second approach is to obtain a value of 6 that minimizes a distance between probability measures rather than the distance of the moment conditions from zero. This data driven approach selects from the set of distributions that satisfy exactly the moment conditions a probability measure Pn closest to the

136

empirical measure Fn defined by the Cressie and Read criterion DP{Fn,Pn)

6

power divergence

= -^-j)YjPi[{nPiy-l],

(6)

where p is a fixed scalar parameter which determines the shape of the criterion function. Cressie and Read 6 proposed the family of power divergence statistics as goodness-of-fit tests. Here, we use the Cressie-Read divergence criterion for estimation purposes. The estimator is defined as the solution to minpne*,eeeDp(Fn,Pn) subject to E[g(x,6)\Pn]

(7)

= / g(x,6)dPn

= 0.

(8)

This form of the analogy principle maps the empirical distribution function onto the space of feasible distribution functions and chooses the probability measure that is most likely to have generated the observed data, subject to the moment conditions (Manski 2 0 ) . The solution to the above constrained optimization problem is a straightforward application of the Lagrange multiplier principle. This framework embeds several interesting special cases (see Kitamura and Stutzer 17 ; and Imbens, Spady and Johnson 1 6 ). The first two cases can also be interpreted as discrete versions of the forward and backward KullbackLeibler discrepancy between the empirical measure and Pn. If we let p approach 0, the estimator is the solution to the problem 2 " min — > In npi p,e

(9)

n *-^

n

n

subject to ^2pig(xi,6)

= 0 and ^ P i = 1.

(10)

This is the empirical likelihood estimator of Owen 24>25>26 and Qin and Lawless 27 obtained as the root of the system of equations

(\Y,U9i{eEL)l(i

+ \'9i{eELj)

\ J'm-\-k^

[h££=i A' ( ^ H J / (i + \'9i(eEL))

J

where A is a vector of Lagrange multipliers on the moment conditions.

137

The (m + k) x 1 parameter vector (6EL, A)' can then be used to compute the vector of probability weights PJ = n f 1 + \'g0EL)) for i = 1,2, ...,n. This gives an efficient estimate of the distribution function F given the set of moment conditions. One drawback of this method is that the form of the loss function resists heavy downweighting of specific data points which may be a problem in the presence of outliers. If p -> — 1, we obtain the maximum entropy, exponential tilting or KLIC (Kullback-Leibler information criterion) estimator of Efron 10 and DiCiccio and Romano 7 and discussed in Kitamura and Stutzer 17 , Imbens 15 and Imbens, Spady and Johnson 1 6 ). It is computed as the minimizer of the KLIC function min 2 V , Pi In npi p

'

(11)

«=i

subject to (10). The maximum entropy estimator solves the system of firstorder conditions \

/^Efeiftfcjexpp'ftfe))

= o.m+k-

l^£r=1A' 26

9gi(0BT)

exp (Xgi(0ET))

Finally, if p -> - 2 , we obtain the Euclidean likelihood estimator of Owen given by the argument that minimizes i

n „2

min P,0

(12)

i)

n i=l

subject to (10). The solution is obtained from \

= o.m+k,

v ^Er=1A'(^#^)[i+A%fe)] where 90EU)

= 90EU)

~ £ E"=i 90 EU)- From the first m equations, A =

" [ £ E ? = i f f i f c ) f f i ( W ] - 1 [^Y:U90EU)] A and j>i = n

E

- (

Pi

_1

1 + \'g~0Eu)

dg0Eu) 89'

• Then, by substituting for

in the last k equations, we get

i n -Y,90EU)-90EU)'

1

ri n -Y,90EU) i=l

=0*.

138

It is interesting to see that this system of first-order conditions is almost identical to (5) for the continuously-updated GMM estimator and very similar to (3) for the iterated GMM estimator. Hence, the Euclidean likelihood estimator can be interpreted as a continuously-updated GMM estimator and an optimally weighted iterated GMM estimator. Note also that the objective functions (9) and (11) implicitly impose the constraint pt > 0 which validates the interpretation of pi as probability weights. For the Euclidean estimator, we either have to inspect the positivity of the probability weights each time or introduce a nonnegativity constraint explicitly into the minimization problem. Owen 26 argues that in small samples, the negativity of the estimated weights may be advantageous for confidence interval construction. 3

Inference in Moment Condition Models

Consider the estimators discussed in Section 2. Let 6Q denote the true value of the parameter vector and suppose that the following regularity conditions are satisfied. Assumption Al. Assume that Wn A- W, where W is a nonstochastic symmetric positive definite matrix; g(xi,9) is continuous in 9; E [sup f l e e |<7(£;,#)|] < oo; supj E [g(xi,0)g(xi, 9)'] < oo for all 9 and 0 is a compact subset of Rk. Assumption A2. There is a unique 9Q such that E[g(xi,90)] E[g(xi,0)] ^ 0 for all 9^90eQ.

= 0 and

Assumption A3. Assume that M = E (d9{Z'/o))

da(

continuous in 9 and E 0o,

su

P0eAT(0o)

is of full rank k;

ee1' \\ < ° ° for

S0me

^;e)

is

neighborhood of

N(90).

Theorem 1. Under Assumptions

A1-A2,

9p -^ #o as n —• oo

v^(£p-0o)4iv(o,n), 1

where Cl = (M'V^M)Proof. See Imbens

1S

and V = E (g(Xi, 90)g(Xi, 90)'). , Qin and Lawless 2 7 , and Newey and Smith

22

.

Theorem 2. Let 9 = (a,/3)', where a € 0 i and j3 € 02 are p x 1 and (k — p) x 1 vectors, respectively. Then, under Assumptions A1-A3,

139 (i) test for overidentifying

restrictions

nDp{9p) 4 x\m-k) (ii) test ofH0:6

as n ->• oo

= 60

n [DP{00) - Dp(9p) ->• x\k) as n -> oo (iii) test for a subset a with Ho : a = a 0 n \Dp(ao,0~p) - Dp(ap,Pp)]

4 x\p) as n ^ oo

where DP{9Q) for p = —2, —1 and 0 is the criterion function defined in (12), (11) and (9), 9p = (ap,j3p) is the unrestricted nonparametric likelihood estimator that minimizes Dp(6) and /3P is the minimizer of Dp(a,j3) subject to a = aoProof. See Imbens

15

, Qin and Lawless

27

, and Newey and Smith

22

.

The results in Theorems 1 and 2 show that we can conduct asymptotically valid inference such as testing for overidentifying restrictions and constructing confidence intervals by inverting the \ 2 acceptance region of the criterion test. The IOOT/% confidence set for the parameter of interest 8 is then given by the set of values of 9 satisfying Cn{x) = {9eQ: h

Dp{9) < qn},

where qn is the lQQrf quantile of the distribution of Dp(9). Equivalently, Cr,(x) = {9 G © : x e A(9)}, where A{9) is the acceptance region of the test Dp{9). The endpoints of the confidence set are the infimum and the supremum over Cv{x), respectively. In particular, the two-sided, equal-tailed confidence interval with nominal coverage n is given by Cn(x) — [9L,9U], where the confidence limits are defined to satisfy 6L — inf{0 6 0 : Vr(Dp(9) < q^Ho) > n} and 9V = sup{9 G 6 : Pr(Dp(9) < qv\H0) > T)}For the weak instrument case, Stock and Wright 29 parameterized the moment condition as a function of the sample size and developed an alternative limiting theory which yields a better approximation to the finite-sample distributions of the estimator and corresponding test statistics. In particular, Stock and Wright 29 replace Assumption 2 with the assumption that E[g(xi,9)] — n _ 1 / 2 m(0) uniformly in 8 G 0 , where m(9) is continuous in 9 and bounded on 0 with m(0o) = 0. Under this assumption, the GMM and nonparametric likelihood estimators are no longer consistent (9P—9Q — Op(l))

140

although the x2 asymptotic approximation for the distributions of the test for overidentifying restrictions and a test of a subvector of weakly identifiable parameters is still valid. 4

Monte Carlo Study

The poor small sample performance of the two-step GMM estimator in linear homoscedastic instrumental variables models with weakly identified parameters has been well documented in Nelson and Startz 21 , Maddala and Jeong 19 , Bound, Jaeger and Baker 4 and Staiger and Stock 28 among others. In this section, we assess the robustness of the nonparametric likelihood estimators in the presence of weak instruments. The structure of the Monte Carlo experiment is similar to the one considered by Angrist, Imbens and Krueger 3 . It is designed to study the finite sample bias of the different estimators and the size properties of hypothesis tests and the test for overidentifying restrictions with a large number of irrelevant instruments. The data are generated from the model Vi = #0 + 6iXi + e»,

(13) z

u

Xi - 70 + L j = l 7j ij + i> where z ~ N(0,I),

(fy

= chol{T,)^ fc ~ iid(0,I),

S = f ^ j j ^fj

,

c/io/(£) denotes Cholesky decomposition of S, 6Q = 0, #i = 1, 70 = 0, 71 = 0.15 and 7/ = 0 for I = 2, ...,m - 1. The optimal two-step GMM estimator in this setting is asymptotically equivalent to the two-stage least squares (2SLS) estimator with Wn = (Z'Z)'1

e2steP = {x'pzxyl

(x'pzy),

where X = (l,x), Z = (l,z) and Pz = Z{Z'Z)~lZ. We also consider the limited information maximum likelihood (LIML) estimator which is given by OLIML

= (X'(I - kMz)Xyl

(X'(I -

kMz)y),

where Mz = I — Pz, k is the smallest characteristic root of (Y Y)(Y MzY)~l and Y = (y X). The confidence intervals for the OLS, 2SLS and LIML estimators are constructed by inverting the Wald test using the x 2 critical values. The confidence intervals for the empirical likelihood (EL) and the Euclidean likelihood

141

(Euclid) estimators are obtained by inverting the \ 2 acceptance regions of the corresponding criterion-based tests discussed above. The results for the exponential tilting (KLIC) estimator are very similar to the results for the EL estimator and are not reported. The number of Monte Carlo replications is 10,000. Table 1. Monte Carlo results for model (13) with T=500 and Gaussian errors. estimator

quantiles around ((?i — 0i) 0.25 0.50 0.75 m — k = l , 7 i = -15 72 = 0 0.6991 0.7154 OLS 0.7336 0.7528 -0.1826 -0.0693 2SLS 0.0348 0.1224 LIML -0.2297 -0.1028 0.0165 0.1184 EL -0.2362 -0.1120 0.0005 0.0940 Euclid -0.2354 -0.1124 0.0011 0.0939 m — k —5,7i = .15 72 = - = 76 = 0 OLS 0.6987 0.7157 0.7339 0.7526 2SLS -0.1098 -0.0148 0.0830 0.1640 LIML -0.2260 -0.0975 0.0193 0.1190 EL -0.2427 -0.1175 -0.0011 0.0959 -0.2402 -0.1179 -0.0022 0.0951 Euclid m — k —10,7i = .15,72 = .. = 7 n = 0 0.7149 OLS 0.6985 0.7338 0.7520 0.1484 2SLS -0.0145 0.0663 0.2220 LIML -0.2291 -0.1020 0.0186 0.1253 -0.2394 -0.1154 -0.0002 0.1009 EL Euclid -0.2408 -0.1161 -0.0010 0.1002 0.10

0.90

test for OIR

coverage rate of CI

0.7695 0.1919 0.2045 0.1655 0.1656

0.1147 0.1219 0.1055 0.1019

0.0000 0.8809 0.8592 0.8909 0.8927

0.7692 0.2297 0.2037 0.1732 0.1720

0.1443 0.1178 0.1118 0.1039

0.0000 0.8080 0.8617 0.8791 0.8830

0.7690 0.2822 0.2104 0.1769 0.1770

0.1837 0.1132 0.1256 0.1059

0.0000 0.6287 0.8470 0.8745 0.8854

First, we assess the effect of increasing the number of redundant moment restrictions on the magnitude of the bias of the estimators and the size properties of the corresponding criterion tests. Tables 1 reports the 0.10, 0.25, 0.50, 0.75 and 0.90 quantiles of the distribution of (#i — 9\) as well as the empirical size of the test for overidentifying restrictions (OIR) with nominal level 0.1 and the coverage properties of the 90% confidence intervals for 6\ in a model with Gaussian errors. Newey and Smith 22 showed that in model (13) with symmetric errors bias(dnML) = bias(8EL) = bias(8Eu) — —fi/n> where S = Claex/a^. Thus, the LIML, EL and Euclidean estimators are higher-order asymptotically

142

equivalent and their bias does not depend on the number of instruments. By contrast, bias(Q2SteP) = (m — 2)6jn which increases linearly with the number of instruments. To investigate the finite-sample sensitivity of the bias of the estimators with respect to the number of irrelevant instruments, we consider three cases of overidentification: m — 1 = 2 (1 overidentifying restriction), m — 1 = 6 (5 overidentifying restrictions) and m — 1 = 11 (10 overidentifying restrictions). Tables 1 contains the results for a sample size of 500 which is commonly encountered in economic applications. As expected, the OLS estimator is severely upward biased. The 2SLS is slightly biased when m — k = 1, but its bias starts to approach the bias of the OLS estimator as m increases. Also, the size properties of the test based on the 2SLS deteriorate significantly as the number of instruments gets large. The magnitude of the bias of the LIML estimator is small and insensitive to the degree of overidentification of the model which is consistent with the theoretical results. The bias of the nonparametric likelihood methods is negligible and it is practically unchanged as the number of the overidentifying restrictions increases. The confidence intervals based on the nonparametric likelihood estimators slightly undercover with the coverage rate of the Euclidean likelihood being closest to the nominal level. Similar results were obtained for sample size T = 150 but these results are not reported due to space limitations. It is interesting to note that the dominance of the Euclidean over the EL estimator in terms of coverage rates is more pronounced for the smaller sample size. This requires further theoretical investigation of the higher-order properties of these tests using Edgeworth expansions. The higher-order asymptotic equivalence of the LIML, EL and Euclidean estimators derived by Newey and Smith 22 is valid only for models with symmetric errors. Newey and Smith 23 show that all members of the class of nonparametric likelihood estimators except EL have an additional bias term coming from the estimation of the variance-covariance matrix fi. To investigate the sensitivity of the results to fat tailed and asymmetric distributions, we also report results from weakly identified models with i-distributed errors with 4 degrees of freedom and x2 -distributed errors with 1 degree of freedom. In addition, we vary the correlation of the endogenous explanatory variable with the instruments. The simulation results in Table 2 show that the nonparametric likelihood estimators provide reliable inference in the presence of weak instruments regardless of the distribution of the errors. Similar to the results in Table 1, the Euclidean estimator dominates in terms of coverage rate with empiri-

143

cal levels within 3 percentage points from the nominal level. Although the bias of the empirical and Euclidean likelihood estimators could be significant for asymmetric errors (for instance, the bottom left corner of Table 2), it is still considerably smaller than the 2SLS and LIML estimators. In summary, the nonparametric likelihood estimators appear to possess good finite-sample properties in overidentified models with weak instruments. Table 2. Monte Carlo results for model (13) with T = 500 and m - k = 1. ^f\

estimator

=

.05,^ 2

=

-02

7i = .05,72 = .05

7! = .05,72 = -1

median

test

cov.rate

median

test

cov.rate

median

test

cov.rate

bias

OIR

of C I

bias

OIR

of C I

bias

OIR

of CI

6~w(o,J) OLS 2SLS LIML EL Euclid

.7909 .0000 .1300 .1352 .8182 .0458 .0995 .8552 .0149 .0859 .8946 .0140 .0852 .8946

.7844 .0877 .0232 .0085 .0077

.0000 .1266 .8405 .1022 .8645 .0967 .8893 .0963 .8903

.7621 .0285 .0286 -.0040 -.0037

.0000 .1075 .8854 .1319 .8462 .0957 .8891 .0950 .8903

.7900 .0000 .1332 .1390 .8190 .0443 .1006 .8502 .0105 .0930 .8842 .0083 .0890 .8943

.7841 .0799 .0184 .0018 .0009

.1261 .1108 .1010 .0957

.0000 .8479 .8715 .8836 .8924

.7602 .0333 .0337 .0031 .0026

.0000 .1124 .8870 .1396 .8344 .1064 .8825 .1016 .8907

.7922 .1595 .1008 .0318 .0287

.1444 .1693 .0955 .0882

.0000 .8043 .7799 .8644 .8776

.7795 .0618 .0333 .0032 .0027

.1168 .1463 .1019 .0940

6 ~ *(4) OLS 2SLS LIML EL Euclid

&~Xa(l) OLS 2SLS LIML EL Euclid

.7957 .2661 .2136 .1126 .1027

.1527 .1720 .0893 .0825

.0000 .7453 .7314 .8545 .8702

.0000 .8559 .8495 .8757 .8908

The finite-sample properties of the constructed confidence intervals for the nonparametric likelihood methods can be further improved by bootstrap methods. For the efficient bootstrap suggested by Brown and Newey 5 and Hall and Presnell 12 , the data can be resampled using the implied probability weights pi from the estimation problem rather than the empirical measure (pi = n~1 for all i) as in the conventional bootstrap. Also, the asymptotic validity of the conventional bootstrap requires explicit recentering of the moment conditions (Hall and Horowitz n ) whereas for the efficient bootstrap the moment conditions, evaluated at the true parameter vector, are centered at zero by construction.

144

5

Empirical Illustration: Return to Education

Estimating the return to education is of central interest to labour economists. It shows the predicted percentage increase in wage for an additional year of education. Since education is believed to be endogenous, Angrist and Krueger 2 suggested the quarter of birth as an instrument for education. However, Bound, Jaeger and Baker 4 challenged the results obtained by Angrist and Krueger 2 arguing that the two-step GMM could be severely biased in the presence of a large number of weak instruments. Here we use the Angrist-Krueger data set which consists of a random sample from 1980 census of 329,500 men who were born between 1930 and 1939. See Angrist and Krueger 2 for a detailed description of the data and model specification. Following Angrist and Krueger 2 , 30 instruments are constructed by interacting quarter and year of birth. Then we draw random subsamples of 500 observations from the original sample without replacement. The results in Table 3 are obtained from 5,000 repetitions and report the median of the parameter estimates and their corresponding standard errors. Table 3. Estimation results for the return of education. EL Euclid OLS 2SLS parameter estimate 0.0708 0.0714 0.0723 0.0653 standard error 0.0087 0.0437 0.0404 0.0373 The estimated return to schooling for all methods is in the range of 6.5% and 7.3%. This is a bit surprising since the weak instruments employed in the estimation are expected to produce a large upward bias in the OLS and 2SLS estimates. This does not seem to be the case and all the estimates do not appear significantly different from one another. It is also interesting to note the smaller standard errors for the nonparametric likelihood estimators compared to the two-step GMM estimator. 6

Concluding Remarks

This paper shows the robustness of nonparametric likelihood estimators of moment condition models to the presence of weak instruments and nonnormal errors. The computational procedure does not involve any explicit bias correction or estimation of variance-covariance matrices. The confidence intervals are obtained directly from the criterion function by inverting its asymptotic acceptance region.

145

One interesting finding that emerges from the study is the existence of noticeable differences in the coverage properties within the class of generalized empirical likelihood estimators. Since the criterion-based test statistics are asymptotically equivalent, higher-order expansions are necessary to appraise the statistical significance of these results. Acknowledgments I would like to thank Gordon Fisher (the Associate Editor) and an anonymous referee for helpful comments and suggestions. Financial support from the FRDP of Concordia University is gratefully acknowledged. References 1. J.G. Altonji and L.M. Segal, Journal of Business and Economic Statistics 14, 353 (1996). 2. J.D. Angrist and A.B. Krueger, Quarterly Journal of Economics 106, 979 (1991). 3. J.D. Angrist, G.W. Imbens and A.B. Krueger, Journal of Applied Econometrics 14, 57 (1999). 4. J. Bound, D. Jaeger and R. Baker, Journal of American Statistical Association 90, 443 (1995). 5. B.W. Brown and W.K. Newey, GMM, Efficient bootstrapping and improved m/erence(Mimeo, Department of Economics, Rice University, 2001). 6. N. Cressie and T. Read, Journal of Royal Statistical Society B 46, 440 (1984). 7. T. DiCiccio and J. Romano, International Statistical Review 58, 59 (1990). 8. S.G. Donald and W.K. Newey, Economics Letters 67, 239 (2000). 9. J.-M. Dufour, Econometrica 65, 1365 (1997). 10. B. Efron, Canadian Journal of Statistics 9, 139 (1981). 11. P. Hall and J.L. Horowitz, Econometrica 64, 891 (1996). 12. P. Hall and B. Presnell, Journal of the Royal Statistical Society B 61, 143 (1999). 13. L.P. Hansen, Econometrica 50, 1029 (1982). 14. L.P. Hansen, J. Heaton and A. Yaron, Journal of Business and Economic Statistics 14, 262 (1996). 15. G.W. Imbens, Review of Economic Studies 64, 359 (1997). 16. G.W. Imbens, R.H. Spady and P. Johnson, Econometrica 66, 333 (1998).

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17. 18. 19. 20. 21. 22. 23.

24. 25. 26. 27. 28. 29.

Y. Kitamura and M. Stutzer, Econometrica 65, 861 (1997). N.R. Kocherlakota, Journal of Monetary Economics 26, 285 (1990). G.S. Maddala and J. Jeong, Econometrica 60, 181 (1992). C.F. Manski, Analog Estimation Methods in Econometrics (Chapman and Hall, New York, 1988). C.R. Nelson and R. Startz, Econometrica 58, 967 (1990). W.K. Newey and R.J. Smith, Asymptotic bias and equivalence of GMM and GEL estimators (Mimeo, Department of Economics, MIT, 2000). W.K. Newey and R.J. Smith, Higher order properties of GMM and generalized empirical likelihood estimators (Mimeo, Department of Economics, MIT, 2001). A. Owen, Biometrika 75, 237 (1988). A. Owen, Annals of Statistics 18, 90 (1990). A. Owen, Annals of Statistics 19, 1725 (1991). J. Qin and J. Lawless, Annals of Statistics 22, 300 (1994). D. Staiger and J.H. Stock, Econometrica 65, 557 (1997). J.H. Stock and J. Wright, Econometrica 68, 1055 (2000).

P R O B A B I L I T Y OF C O R R E C T SELECTION OF G A M M A V E R S U S G E OR WEIBULL V E R S U S GE B A S E D O N LIKELIHOOD RATIO STATISTIC RAMESHWAR D. GUPTA Department of Applied Statistics and Computer Science The University of New Brunswick, Saint John Canada, E2L 4L5 E-mail: [email protected]

DEBASIS KUNDU AND ANUBHAV MANGLICK Department of Mathematics Indian Institute of Technology Kanpur Pin 208016 India. E-mail: [email protected] This paper proposes the use of likelihood ratio statistic in choosing between gamma and GE models or between Weibull and GE models. Probability of correct selections are obtained using Monte Carlo simulations for various sample sizes and for various model parameters. Simulation results indicate that it is easier to distinguish between GE and Weibull models than between GE and gamma models. Two real life data sets are analyzed. Interestingly, in both cases although in the literature gamma or Weibull model was used but based on the maximum likelihood values we select GE as the 'best' fitted model among these three distributions.

1

Introduction

Recently a new two-parameter distribution named as Generalized Exponential (GE) distribution or Exponentiated Exponential distribution has been introduced and studied quite extensively by two of the authors Gupta and Kundu 6,7,8,9,10 rpj^ Qg f a m iiy h ^ the following distribution function FGE(x;a,\)

= (l-e-Xx)a;

a,\>0,

density function fGE(x;a,\)=a\{l-e-Xx)a-1e-Xx;

a,X>0,

survival function SGB(x;a,\)

= l-(l-e-Xx)a;

a,X>0,

and the hazard function . , ., aA(l - e-x*)"-1e-Xx hGE(x;a,\) = 1_(1_e_Ax)a

147

^

. n a,A>0.

(1)

148

Here a and A are the shape and scale parameters respectively. It is observed in Gupta and Kundu 6 ' 7 that the two-parameter GE distribution can be used quite effectively in analyzing many lifetime data particularly in place of twoparameter gamma or two-parameter Weibull distribution. The two-parameter GE distribution can have increasing, decreasing and constant failure rates depending on the shape parameter (Gupta and Kundu 6 ) . It is also observed by Gupta and Kundu 9 that the behavior of the two-parameter GE distribution is very much similar to a gamma distribution in many respects and in many cases the two-parameter GE model provides a better fit than the two-parameter gamma model or the two-parameter Weibull model in terms of maximum likelihood or minimum chi-square. Therefore, it is quite important to choose between a Weibull and GE models or between a gamma and GE models to analyze real life data sets. It is found to be very difficult to discriminate between these three models because all these three models are quite flexible and they overlap with each other in the sense that exponential distribution is a special case to all of them. Although, these three models may provide similar data fit for moderate sample sizes but it is still desirable to select the correct or more nearly correct model, since the inferences based on the model will often involve tail probabilities where the affect of the model assumption will be more critical. Therefore, even if large sample sizes are not available, it is still important to make best possible decision based on whatever data are available. In this paper we propose to use the logarithm of the ratio of the maximum likelihood functions in choosing two overlapping distributions. The idea was originally proposed by Cox 3 | 4 in discriminating between two models and Bain and Englehardt 1 used it in choosing between gamma and Weibull models. It is observed that the probability of correct selection (PCS) depends only on the shape parameter of the distribution from which the data are coming. Since it is not possible to obtain the exact distributions of the likelihood ratio statistics, we obtain PCSs by using extensive Monte Carlo simulations for various sample sizes and for various model parameters. We use two real data sets to illustrate how the proposed methods can be used in practice. Rest of the paper is organized as follows. In section 2, we provide the selection method based on likelihood ratio statistic. The PCSs are presented in section 3. We analyze two real data sets in section 4 and finally we conclude the paper in section 5.

149

2

Likelihood Ratio Statistic

Suppose X\,..., Xn are independent and identically distributed random variables from any one of the three distributions. The density function of a GE random variable with the shape parameter a and the scale parameter A will be denoted by (1). For both gamma and Weibull distributions we denote the shape parameter by /? and the scale parameter by 9. The density function of a gamma random variable with the shape parameter (3 and the scale parameter 9 will be denoted by JGA{X) as /GA(X)

= j ^ ^ e - ' * ;

x,/3,9>

0,

(2)

and similarly the density function of a Weibull random variable with the shape parameter parameter (3 and the scale parameter 9 will be denoted by fWE(x)

=

P0
Let us define n

n

LGE(a, A) = J J fGE{xi),

LGA(/3,9) = J J fGA(xi)

i=l

i=l

and n

=

LWE(P,0)

Y[fwE(xi).

Now first consider choosing between gamma and GE models. The natural logarithm of the likelihood ratio statistic T\ = ln(Li), where _

LGE(a,X)

1_

LGA0,9)

and 7\ = n \n(a\X)

- p ln(X0) - ^ . ^ + ln(r(/?)) - X(X - 9) a

(3)

Here X = ^ Y^Ji=\ -^»> -^ = (I"Ii=i -^i)" • Moreover, & and A are maximum likelihood estimators (MLEs) of a and A if the data are assumed to come from a GE distribution and in this case (Gupta and Kundu 7 ) they have the following relation a—

ElLiln(l-e-^)

150

Similarly /? and 9 are MLEs of /3 and 9 if the data are assumed to come from a gamma distribution and in this case they satisfy the following equation

'4The proposed procedure is as follows. Choose a GE model if Ti > 0, otherwise choose a gamma model. It is clear from (3) that if the data are coming from a GE population, the distribution of T\ is free of A and depends only on a and similarly if the data are coming from a gamma population, the distribution of T\ is independent of 6 and it depends only on /?. The above assertions can be proved very easily. First let us assume that the data are coming from GE(a, A). It implies that the distribution of XXi is independent of A. Moreover it follows from the Theorem 7.8.5 of Bain and Englehardt 2 that the distributions of j and j are independent of A. Since the distributions of a and $ are independent of A, therefore the result follows immediately from the expression of Ti. Similar argument can be given when the data are coming from a gamma distribution then the distribution of T\ is independent of 9 and depends only on /?. It implies that in the first case PCS is free from A and depends only on a, whereas in the second case PCS is free from 9 and depends only on 0. Now we consider the likelihood ratio statistic for discriminating between GE and Weibull models. Similarly as before, the natural logarithm of the likelihood ratio T2 = ln(L 2 ), where _

LGE{OIA)

~

LWE0,§)

2

and To. = n ln(^\-^l-XX-pin(9X)

+ l

(4)

Here X and X are same as defined before and a and A are MLEs of a and A for GE distribution and similarly /? and 9 are MLEs of /? and 9 for Weibull distribution. In the case of Weibull distribution, 9 and J3 satisfy the following relation 9=(

n

En

yf.

We use the similar discrimination procedure as before, i.e. if T2 > 0, choose GE distribution, otherwise choose Weibull distribution. In this case also it

151 Table 1. Probability of correct selection between GE and gamma distributions and when the data are generated from a GE distribution

n | a —>

20 40 60 80 100 200

0.50 0.6610 0.6342 0.6282 0.6134 0.6134 0.6033

0.75 0.5798 0.5870 0.5847 0.5822 0.5813 0.5760

1.0

2.0

4.0

8.0

0.4942 0.5004 0.5039 0.5047 0.5059 0.4945

0.4633 0.4812 0.5100 0.5194 0.5321 0.5699

0.5081 0.5563 0.5841 0.6059 0.6242 0.6947

0.5503 0.6155 0.6472 0.6740 0.7045 0.7860

16.0 0.6012 0.6566 0.7031 0.7421 0.7641 0.8601

can be shown as before that if the data come from a GE population then the distribution of Ti depends on a and independent of A and if the data come from a Weibull population, then the distribution of T
3

Probability of Correct Selections

In this section we use Monte Carlo simulations to compute the PCSs for different shape parameters and for different sample sizes. We consider different shape parameters namely 0.50, 0.75, 1.0, 2.0, 4.0, 8.0, 16.0 and also different sample sizes namely 20, 40, 60, 80, 100, 200. All the probabilities are calculated based on 10,000 replications. The exact details are provided below: Case 1: Between GE and Gamma First we generate a sample of size n from a GE(a, 1). From the given sample we obtain MLEs of a and A (GE parameters), similarly we obtain MLEs of /? and 9 (gamma parameters). We compute T\ and observe whether it is negative or positive. We replicate the process 10,000 times and compute the percentage of times T\ is positive and that provides the probability of the correct selection. The same way we estimate the PCS between gamma and GE models when the data are coming from a Gamma(/3, 1). We generate the sample from a Gamma(/3, 1) and obtain the MLEs of a, A, /3 and 9. In this case we compute the percentage of times Ti is negative out of 10,000 replications. Results are reported in Tables 1 and 2.

152 Table 2. Probability of correct selection between GE and gamma distributions when the data are generated from a gamma distribution

n i p -> 20 40 60 80 100 200

0.50 0.3655 0.4086 0.4370 0.4555 0.4519 0.5039

0.75 0.4246 0.4317 0.4398 0.4498 0.4597 0.4855

1.0

2.0

4.0

8.0

0.4977 0.5104 0.4999 0.4964 0.4998 0.4991

0.5829 0.5948 0.5942 0.6051 0.6081 0.6319

0.6104 0.6281 0.6607 0.6745 0.6943 0.7534

0.6347 0.6900 0.7197 0.7544 0.7775 0.8598

16.0 0.6653 0.7365 0.7849 0.8276 0.8542 0.9317

Table 3. Probability of correct selection between GE and Weibull distributions when the data are Generated from a GE distribution.

n 4- a —> 0.50 20 0.6610 40 0.7258 60 0.7663 80 0.7856 100 0.8128 200 0.8849

0.75 0.5643 0.5982 0.6157 0.6309 0.6430 0.7012

1.0

2.0

4.0

8.0

0.5025 0.4976 0.4849 0.5042 0.5097 0.4933

0.5731 0.6316 0.6829 0.7139 0.7382 0.8295

0.6675 0.7520 0.8154 0.8582 0.8825 0.9614

0.7283 0.8357 0.8929 0.9224 0.9496 0.9911

16.0 0.7699 0.8811 0.9301 0.9582 0.9752 0.9981

Case 2: Between GE and Weibull In this case first we compute the PCS between GE and Weibull distributions and when the data are coming from a GE distribution. We generate a sample of size n from a GE(a, 1) and compute MLEs of a, A (GE parameters) and /3, 6 (Weibull parameters). Replicate the process 10,000 times and observe the percentage of times T^ is positive. Exactly the same way, we compute the PCSs between GE and Weibull distributions and when the data are coming from a Weibull distribution. The results are reported in Tables 3 and 4 a. Some of the points are quite clear from the Tables 1,2,3 and 4. First of all in all four cases when a = 1.0, the PCS is close to .5 as it should be for all sample sizes. Because when a = 1.0, all three distributions become exponential distribution therefore, between any two distributions the probability of choosing any particular one is 0.5. It is also observed that in all cases (except for a = 0.50 or 0.75 in Table 1) as sample size n increases the PCS increases "When /3 = 4, the Weibull distribution becomes almost symmetric. In this case it is too time consuming to compute MLEs of the GE parameters. In Table 4, for /3 = 4, the results are based on 1000 replications.

153 Table 4. Probability of correct selection between GE and Weibull distributions when the data are generated from a Weibull distribution.

n I 0 -> 20 40 60 80 100 200

0.50 0.6655 0.7159 0.7768 0.8233 0.8653 0.9492

0.75 0.5155 0.5844 0.6169 0.6457 0.6700 0.7536

1.0

2.0

4.0

8.0

0.5037 0.5045 0.5046 0.5069 0.5047 0.5085

0.7061 0.7823 0.8395 0.8757 0.9006 0.9660

0.8142 0.9185 0.9573 0.9768 0.9874 0.9992

0.8738 0.9624 0.9868 0.9947 0.9979 1.0000

16.0 1.0000 1.0000 1.0000 1.0000 1.0000 1.0000

for all values of a ^ 1. Moreover as |Q — 1| increases the PCS increases from 0.5 in most of the cases. Surprisingly in Table 1, when the sample size 20 or 40 and the a = 2.0, the PCSs are less than .5. It indicates that when the data come from GE and the shape parameter is greater than one but not very far from one then the likelihood ratio statistic can not distinguish properly whether the data are coming from a GE distribution or from a gamma distribution if the sample size is not very large. Similar pattern is also observed in Table 2, when the shape parameter is 0.50 or 0.75. It indicates that for certain ranges of the shape parameter it is really very difficult to distinguish between gamma and GE distributions. In Table 1, when a = 0.50 or 0.75, the PCSs decrease as n increases. In Table 2, when a = 0.50 and 0.75, the PCS are less than 0.5. Both these findings are quite counter intuitive and we can not justify them. We feel it might be due to the estimation procedures. It is known that when the shape parameters is less than one then the parameter estimations are quite difficult for both GE and gamma distributions. The likelihood functions are very flat and therefore the maximum likelihood estimators are quite unstable. The results might be the reflection of that. One of the referees has mentioned that it might be due to discretization of the cumulative distribution function. Now comparing the results of Tables 1 and 2, it clear that the PCSs of Table 2 are more than the corresponding PCSs of Table 1 when the shape parameter is more than one. It clearly indicates that for the shape parameter greater than one, the likelihood ratio statistic can distinguish better between gamma and GE distributions if the data are coming from a gamma distribution than vice verse. It can be justified as follows. As the shape parameter increases the gamma density becomes symmetric but the GE density remains skewed as the shape parameter goes to infinity. Therefore, if the data are coming from a gamma distribution and the shape parameter is large then

154

naturally PCS becomes higher. Comparing the results between Tables 3 and 4, it is also observed that PCSs of Table 4 are more than the corresponding PCSs of Table 3. Therefore, between GE and Weibull distributions also the likelihood ratio statistic can distinguish better if the data are coming from a Weibull distribution than the other way. The same reason as above holds here also. Finally comparing the results between Tables 1, 2 and Tables 3, 4, it is clear that using the likelihood ratio statistic it is much easier to distinguish between Weibull and GE models than between gamma and GE models. It might be due to the fact that the Weibull density goes to symmetry much faster than the gamma density as the shape parameter increases. We believe that it should be true for any other statistics also and more work is needed in this direction. 4

D a t a Analysis

In this section we apply the above procedure in two data sets. Both these data sets have been used by several authors in the literature. D a t a Set 1: Lieblein and Zelen n provided the following sample of size n = 23 to illustrate the use of Weibull model. These data indicate the endurance in millions of revolutions of deep-groove ball bearings. Thoman, Bain and Antle 12 mentioned that Weibull distribution should be used where as Bain and Englehardt * proposed to use gamma distribution. The data are as follows: 17.88, 28.92, 33.00, 41.52, 42.12, 45.60, 48.80, 51.84, 51.96, 54.12, 55.56, 67.80, 68.64, 68.64, 68.88^84.12, 93.12, 98.64, 105.12, 105.84, 127.92, 128.04, 173.40. For this data X = 72.12 and X = 63.46. The MLEs of the GE parameters are a = 5.2825, A = 0.0323, gamma parameters are ft = 4.0309, 9 = 0.0558 and Weibull parameters are /3 = 2.1033, § - 0.0122. Also ln{LGE(a,X) = -112.9762, ln{LGA(/3,§) = -113.0272 and ln(LWE0J) = -113.6886. Therefore, Tx = 113.0272-112.9762 = .0510 and T2 = 113.6886-112.9762 = 0.7124. Since both T\ and T? are positive, therefore GE is preferable than gamma or Weibull distributions based on the likelihood ratio statistic. Moreover, from the tables it can be said that the PCS between gamma and GE is between 51% to 55% and the PCS between Weibull and GE is between 67%-73%. D a t a Set 2: The following data represents the relief times of 20 patients receiving an analgesic. The data is obtained from Gross and Clark 5 (page; 105), where they fitted gamma distribution. The data are as follows: 1.1, 1.4, 1.3, 1.7, 1.9, 1.8, 1.6, 2.2, 1.7, 2.7, 4.1, 1.8, 1.5, 1.2, 1.4, 3.0, 1.7, 2.3, 1.6, 2.0. For this data set X = 1.90 and X = 1.80. The MLEs of the GE parameters are as follows: a = 36.6437, A = 2.2348 and ln(LGE(a,X)) = -16.2605.

155

The MLEs of the gamma parameters are (1 = 9.6630 and § = 5.0864 and \n(LGA0,6)) - -17.8186. Similarly we obtain the MLEs of the Weibull parameters as 0 = 2.7875 and 6 = 0.4695 and \n{LWE{P,6)) = -20.5864. Therefore, Ti = -16.2605 + 17.8186= 1.5581 and T2 = -16.2605 + 20.5864 = 4.3259. Since T\ and T-i both are positive therefore in both cases we prefer GE models. The PCS between gamma and GE is more than 60% and the PCS between Weibull and GE is more than 77%. 5

Conclusions

In this paper we consider the likelihood ratio statistics to choose between GE and gamma model or between GE and Weibull models. It is very easy to use in practice. We could not obtain the exact distributions of the likelihood ratio statistics so we use Monte Carlo simulations to compute the probability of correct selections. The likelihood ratio statistics depend only on the shape parameters. It is observed that as the shape parameter deviates from one it becomes easier usually to distinguish between two distributions. It is also observed that it is much easier to distinguish between GE and Weibull distributions than between GE and gamma distributions. Two data sets are analyzed. Interestingly, in both cases although in the literature gamma or Weibull model was used but we select GE as the 'best' fitted model based on the maximum likelihood values. Another interesting question is to determine the minimum sample size n to discriminate between two distribution functions (either GE and gamma or GE and Weibull) for a given PCS. Work is in progress in that direction and it will be published elsewhere. Acknowledgments Part of the work of the first author was supported by a grant from the Natural Sciences and Engineering Research Council. The authors would like to thank two anonymous referees for some constructive suggestions. References 1. L.J. Bain and M. Englehardt, Communications in Statistics Theory and Methods 9, 375 (1980). 2. L.J. Bain and M. Englehardt, Statistical Analysis of Reliability and Lifetesting Models; Theory and Methods (Marcel and Dekker Inc., New York, 1991)

156

3. D.R. Cox, Proceedings of the Fourth Berkeley Symposium in Mathematical Statistics and Probability 5, 105 (1961). 4. D.R. Cox, Journal of the Royal Statistical Society Ser. B 24, 406 (1962). 5. A.J. Gross and V.A. Clark Survival Distributions; Reliability Applications in the Biomedical Sciences (John Wiley and Sons, New York, 1975). 6. R.D. Gupta and D. Kundu, Australian and New Zealand Journal of Statistics 4 1 , 173 (1999). 7. R.D. Gupta and D. Kundu, Biometrical Journal 43, 117 (2001). 8. R.D. Gupta and D. Kundu, Jour. Stat. Comp. Simul 69, 315 (2001). 9. R.D. Gupta and D. Kundu, Closeness of gamma and generalized exponential distributions (Technical Report, Department of Mathematics, Statistics and Computer Science, The University of New Brunswick, New Brunswick, Canada, 2001). 10. R.D. Gupta and D. Kundu, Jour. Appl. Stat. Sc. (Accepted, 2002). 11. J. Lieblein and M. Zelen, Jour. Res. Nat. Bur. Stand. 47, 273 (1956). 12. D.R. Thoman, L.J. Bain and C.J. Antle, Technometrics 11, 445 (1969).

SURVEY M E T H O D O L O G Y FOR S T U D Y I N G S U B S T A N C E U S E P R E V E N T I O N P R O G R A M S IN SCHOOLS S H E L T O N M. J O N E S , B R I A N C. S U T T O N A N D K E R R I E E . B O Y L E

E-mail:

RTI International, P. O. Box 12194, Research Triangle Park, NC 27709-2194, USA [email protected]; [email protected]; [email protected]

According to the 1998 National Household Survey On Drug Abuse, almost 10 percent of youths in the United States between ages 12-17 used illicit drugs SAMSHA (Substance Abuse and Mental Health Services Administration n ) . In addition, 18 percent of 12-17 year olds were current cigarette users, over eight percent were current users of marijuana, and almost one percent of youths were reported to be current users of cocaine. These percentages demonstrate a need for continuous substance use prevention programs. Although such programs have existed in U. S. school systems for some time, their extent and prevalence have been unclear (Ringwalt et al. 1 0 ; Kann et al. 7 ) . The School-Based Substance Use Prevention Programs Study (SSUPPS) is designed to study substance use prevention activities currently available in middle schools. In addition to a school sample component, SSUPPS includes a public school district component for those districts associated with the public school sample. This paper presents survey design methodology for obtaining statistically valid national estimates of substance use prevention programs among the conventional public and private middle schools in the United States. Also, presented is information on sample selection and sample weighting procedures.

1

Introduction

Most researchers tend to agree that the best way to prevent substance use among adolescents is to reduce demand (Dusenbury and Falco 3 ; Eigen and Rowden 4 ). Substance abuse prevention programs for many adolescents are sponsored and maintained by schools with Federal assistance from the United States Safe and Drug-Free Schools and Communities Act (SDFSCA). These programs vary widely, and include both classroom curricula and nonclassroom activities. The School-Based Substance Use Prevention Programs Study (SSUPPS) is designed (1) to determine if these prevention programs differ across schools with middle school grades by school type, enrollment size, urbanicity, and students socio-economic status, and (2) to determine if the established programs reflect evidence of effectiveness provided by prevention research. This paper presents the sample design established to provide descriptive information about the prevalence and characteristics of school-based programs. Probability sampling of middle schools in the U.S. is the basis for our stratified systematic sampling design (see Cochran x ) of both public and

157

158

private schools. For the list of acronyms used, please see the appendix. 2

Defining the Target Population

SSUPPS is a survey of conventional schools with middle school grades and public school districts for the 1998/1999 academic year. Middle schools are defined as conventional schools with one of the following grade combinations: (1) only fifth and sixth grades, (2) only a sixth grade, or (3) a seventh or an eighth grade. Specifically, this study provides information on substance use prevention programs available for students approximately 10-13 years of age. Excluded from the target population are non-conventional schools or educational units (such as special education schools, alternative education schools, public middle schools with fewer than 20 students enrolled, State Department of Education units, charter schools, Bureau of Native-American Affairs units, adult education schools, Department of Defense units, vocational technical schools, and middle schools with no eligible student enrollment for the 1998/1999 academic year). A public school district is an educational unit of authority that operates one or more public schools. 3

Sampling Frame Considerations

A sampling frame is defined as the list or mechanism used to identify population elements for the selection of a sample. There are three conventional sources of sampling frames for school data. The Common Core of Data or CCD (U.S. Department of Education 15 ) is the primary public school data source of the National Center for Education Statistics (NCES). Two other sources are owned by Market Data Retrieval or MDR (Dun & Bradstreet Corporation 2 ) and QED (Quality Education Data, Inc. 9 ). The MDR includes all educational levels from preschool through college in the United States and Canada. Both the CCD and the MDR have proven to be reliable sources of information on public school data (Hamann 5 ) . A single data source was preferred to reduce school multiplicity and therefore eliminate the need to merge multiple files to capture the entire target population. The QED includes information on urbanicity and poverty that can be beneficial for stratification. In addition, QED provides educational data useful for all types of schools: public, Catholic, and non-Catholic schools from pre-kindergarten through 12th grade. Therefore, due to the need to survey both public and private schools, as well as the need for stratification, the QED database was chosen as the sampling frame. It should be noted that since the sampling frame for this study was constructed, there have been important developments in school

159

lists. NCES now maintains, in addition to the CCD, a database of non-public schools by means of the Private School Universe Survey (PSS). Both CCD and PSS are accessible on the NCES website (http://nces.ed.gov). Therefore, if this research was conducted in the future, we would recommend a combination of CCD and PSS. Once finalized, the sampling frame contained a total of 37,841 eligible conventional middle schools (because the classifications of ineligibles are not mutually exclusive, the ineligible counts are not additive). This frame comprised 22,891 public middle schools, 6,031 Catholic schools, and 8,919 non-Catholic private middle schools (Table 1). 3.1

Explicit and Implicit

Stratification

The sampling frame was stratified explicitly and implicitly to control the distribution of the sample and to reduce sampling variation among survey statistics (see Kish 8 , pp. 75-112). Explicit stratification was used to provide domain estimation for public schools at different levels of urbanicity, school size, and poverty index. Urbanicity was defined at three levels provided by QED: urban (area within the central city), suburban (area surrounding the central city), and rural (area outside any metropolitan area). Three levels of size were defined based on the estimated total number of eligible students enrolled in middle school grades: small (fewer than 200 eligible students), medium (200-600 eligible students), and large (more than 600 eligible students) . The poverty index for public schools was determined from the Orshansky percentage, which is the percent of students who fall below the federal governments poverty guidelines. The Orshansky percentage is defined at the district level and is a relative indicator of community wealth/poverty when compared to other school districts. Three poverty index levels were formed: low (schools with an Orshansky percentage 10% or less), medium (11-40 %), and high (greater than 40 %). The various combinations of strata for public schools formed a total of 27 levels of explicit stratification (Table 1). Due to the small population of non-public schools and because of secondary interest, non-public schools were only stratified by Catholic and non-Catholic to assure that the non-public school population received some representation. Additionally, state identification was used to implicitly stratify the public and non-public school sampling frames. 4

Sample Selection

A school sample component was designed to survey a probability sample of teachers responsible for teaching substance use prevention programs at the

160

school. In addition, a separate survey was conducted of Safe and Drug-Free School District Coordinators. The district survey was conducted only among those districts that oversaw schools in the sample. The sample design consisted of selecting a stratified systematic sample of middle schools within each explicit design stratum as described by Cochran J (pp. 226). A sample of schools or teachers and district coordinators was selected to participate in a mail survey on substance use prevention programs at the school or district. The sample design is illustrated in Figure 1. 4-1

The Middle School Sample

The initial middle school sample of 3,430 schools accommodated two data collection waves. Wave 1 consisted of a random subsample of 2,852 school selections for possible participation in the study. The remaining schools comprising Wave 2 were not surveyed due to time constraints relative to data collection. Although strict proportional allocation was not used, adequate representation in each stratum made it possible for the public school sample to cover a broad range of different types of middle schools. The non-public school sample of 472 schools was proportionally allocated to Catholic and other private schools for a sample of 189 Catholics and 283 non-Catholics. 4-2

Under-Coverage or

Under-Representation

When the frame was constructed, QED was one of the most reliable data sources for public and non-public schools. Public schools are updated four times per year, and both public and non-public schools are re-verified every two years. However, the list of non-Catholic private schools is not as complete as that of public schools. This is primarily because of the difficulty in securing information on non-Catholic private schools, since many tend to be small and difficult to track from year to year. Because of this limitation, nonCatholic private schools are under-represented beyond the under-sampling of non-public schools mentioned above. 4-3

Inclusion of Public School Districts

This study primarily surveyed teachers who had knowledge of substance use prevention programs within schools with middle school grades. Due to secondary interest in district level inference, the survey design was expanded to accommodate a comprehensive survey of public school districts. Each district was included in the sample based on the number of schools associated with the district and selected from the stratum.

161

5

Sample Weighting and Results

Sample weights for schools were computed as the inverse of the school sample selection probabilities. The district weights were determined as a function of a multiplicity adjustment estimator (Sirken 1 4 ). Some sample schools and district coordinators refused to participate in the study. This type of nonresponse is referred to as unit nonresponse because information is missing for the entire sampling unit. The initial sampling weights of the responding units were adjusted to compensate for the missing data arising from the nonrespondents. A common weighting procedure referred to as sample-based adjustment cell weighting is described by Jones and Chromy 6 . With the school sample, adjustment cell weighting was employed to partition the school respondents into adjustment cells or weighting classes. Within each class or stratum, responding teachers were weighted up in an attempt to compensate for the nonresponding teachers within each stratum. These respondents weights were adjusted by the sum of the stratum weights for the respondents plus the nonrespondents divided by the sum of the respondents weights. The nonresponse adjusted weights are recommended for use in all weighted statistical analyses with software packages such as, SUDAAN (Shah, Barnwell, and Bieler 13 ) and SAS (SAS Institute Inc. 1 2 ) . 5.1

School Weights

For our design, each school within each explicit stratum was selected with the same probability. Let Nh, denote the number of schools on the frame within stratum h, and nh denote the number of schools selected within stratum h (h = 1,2, ...,29) based on frame information, which occasionally is not accurate. It follows that the school sample selection probability {ITM) a n d the initial sampling weight (whi) for school i within stratum h are respectively given by:
162

37,841 Middle Schools on Sampling Frame

—>

—> 3,430 Middle Schools Selected (Waves 1 & 2) 1

2,380 Public Middle Schools Selected

6,031 Catholic Schools

8,991 Other Private Schools

2,858 Public Schools Selected (Waves 1 & 2)

229 Catholic Schools Selected (Waves 1 & 2)

343 Other Private Schools Selected (Waves 1 & 2)

i —>

i

22,891 Public Schools

2,380 Publics (Wave 1)

Metro Type Urban

Subarban

Rural

Total

1 Orshansky Score2 Low Medium High Total Low Medium High Total Low Medium High Total Low Medium High Total

i 189 Catholics (Wave 1)

I 283 Privates (Wave 1)

Eligible Grade Size3 Total Small Medium Large 82 10 33 39 352 134 178 39 222 119 20 83 655 69 250 336 332 152 138 42 442 202 170 70 167 75 57 35 941 147 429 365 139 14 57 68 468 322 125 21 177 41 130 6 784 41 520 223 553 242 120 191 1,261 369 431 461 566 182 199 185 742 2,380 902 736

1 The initial sample included a subsample of 2,852 schools in Wave 1 and 578 schools in Wave 2. Wave 1 was surveyed, but due to time constraints, wave 2 was not. 2 The Orshansky scores are categorized as: Low - less than or equal to 10%, Medium - 11 to 40%, and High - 4 1 % and higher. 3 Eligible grade sizes are estimates of the total number of students enrolled at the school for all Middle School specific grades (5-8). The size distinctions are: Small - fewer than 200 students, Medium - 200 to fewer than 600 students, and Large - 600 and more students.

Figure 1. T h e S S U P P S S a m p l e

163

5.2 District Weights Since schools were sampled within strata based on school level characteristics, it was possible that a district contained multiple schools across different strata. Thus, it was sometimes impossible to match each district with a unique stratification level. Therefore, district weights were created by expanding or replicating the districtlevel file so that the number of observations equaled the number of sampled eligible schools. These weights were post-stratified to the sampling frame totals by three Orshansky levels (less than 10%, 11% to 40%, and greater than 41%). Finally, the weights for responding districts were adjusted for district level nonresponse. Given the definitions in Section 5.1, let Mj be the number of eligible schools on the frame within district j , and rtihj denote the number of times district j was included within stratum h, then the basic sampling weight (w(p)j) for district j is:

Denoting the number of sample districts in Orshansky level o, (o = 1,2,3) by n0 and the number of eligible schools selected in district j by rij, equation (1) provides the sampling weight (tU(„)j-) for weighted estimation as: w^j = w^j/rij. The post-stratified weight (w?*>.) for district j is: w?V = W(„)j(n 0 / VJ0 W(„)j), where S o w (")j = s u m °f initial district weights for point and variance estimation by Orshansky levels (o = 1,2,3). The non-response adjusted weight (u>(v)j) for district j is :

y; we,vs »?-M = <W I ^y Tw7^ 1 • where TJ o w^f. = sum of post-stratified weights for eligible districts, and y j 0 w^f sum of post-stratified weights for responding districts. 5.3

Sampling

P) =

Results

Table 1 shows various unweighted eligibility and response rates from school sample. Rates are provided for 27 different levels of explicit public school strata plus 2 levelest for non-public schools, including three domain variables of interest (urbanicity, Orshansky percent, and schools size). Individual Orshansky-level eligibility and response rates are also given in Tables 3 and 4. Eligibility rates for both the public and Catholic schools (95.5 and 92.1) proved significantly higher than rates from non-Catholic private schools (71.0). Response rates range from a low of 46.7

164 Table 1. Frame Count with Number Selected, Eligibles, and Respondents by Strata for SSUPPS Frame Count Design S t r a t u m (urbanicity, Orshansky , size )

of

Schools

Number of Selected Schools

Eligible Schools

Respond i n g Schoo s

Number

Percent

Number

Percent

448 53

10 33 39 39 134 178 20 83 119 42 152 138 70 202 170 35 75 57 68 57 14 322 125 21 130 41 6

8 33 38 33 127 172 15 78 114 37 148 135 60 192 165 32 70 56 65 56 14 314 123 20 124 39 5

80.0 100.0 97.4 84.6 94.8 96.6 75.0 94.0 95.8 88.1 97.4 97.8 85.7 95.0 97.1 91.4 93.3 98.2 95.6 98.2 100.0 97.5 98.4 95.2 95.4 95.1 83.3

6 17 32 17 89 133 7 60 82 27 102 99 46 148 121 22 51 39 49 41 11 232 96 14 83 29 3

75.0 51.5 84.2 51.5 70.1 77.3 46.7 76.9 71.9 73.0 68.9 73.3 76.7 77.1 73.3 68.8 72.9 69.6 75.4 73.2 78.6 73.9 78.0 70.0 66.9 74.4 60.0

Urban Suburban Rural

4,816 9,720 8,355

655 941 784

618 895 760

94.4 95.1 96.9

443 655 558

71.7 73.2 73.4

Orshansky (low) Orshansky ( m e d i u m ) Orshansky (high)

4,764 12,700 5,427

553

534

1261

1206

566

533

96.6 95.6 94.2

384 896 376

71.9 74.3 70.5

Size (small) Size (medium) Size (large)

7,603 8,649 6,639

736 902 742

688 866 719

93.5 96.0 96.9

489 633 534

71.1 73.1 74.3

Public School

22,981

2,380

2,273

95.5

1,656

72.9

30 - Catholic 40 - Other Private

6,031 8,919

189 283

174 201

92.1 71.0

129 120

74.1 59.7

Public Schools 1 - urban, low, small 2 - urban, low, medium 3 - urban, low, large 4 - urban, medium, small 5 - urban, m e d i u m , m e d i u m 6 - urban, medium, large 7 - urban, high, small 8 - urban, high, medium 9 - urban, high, large 10 - s u b u r b a n , low, small 11 - s u b u r b a n , low, medium 12 - s u b u r b a n , low, large 13 - s u b u r b a n , medium, small 14 - s u b u r b a n , medium, medium 15 - s u b u r b a n , medium, large 16 - s u b u r b a n , high, small 17 - s u b u r b a n , high, m e d i u m 18 - s u b u r b a n , high, large 19 - rural, low, small 20 - rural, low, medium 21 - rural, low, large 22 - rural, m e d i u m , small 23 - rural, m e d i u m , m e d i u m 24 - rural, medium, large 25 - rural, high, small 26 - rural, high, m e d i u m 27 - rural, high, large

57 236 280 262

1,007 1,333 140 615 886 374

1,381 1,219 776

2,242 1,883 383 830 632 592 505 120

3,579 1,385 233

1,440

Non-Public School

14,950

472

375

79.4

249

66.4

All Schools

37,841

2,852

2,648

92.8

1,905

71.9

T h e Orshansky scores are categorized as: Low - less than or equal to 10%, Medium - 11 to 40%, and High 4 1 % and higher. Eligible grade sizes are estimates of the t o t a l number o students enrolled at school for all middle school specific grades (5-8). T h e size distinctions are: Small - fewer than 200 s t u d e n t s , Medium - 200 to fewer t h a n 600 s t u d e n t s , and Large - 600 and more s t u d e n t s .

165 Table 2. Public School District Level Eligibility and Response Rates Design Stratum (Orshansky) Low Medium High Total

Eligible Districts Number Percent 518 98.5 1,051 98.3 98.1 415 98.3 1,984

Number of Districts Included 526 1,069 423 2,018

Responding Districts Number Percent 403 77.8 81.4 856 80.5 334 80.3 1,593

Table 3. School Eligibility Percentage by School Type School Type Percentage

Public 95.5

Catholic 92.1

Other Private 71.0

Total 92.8

Table 4. School Responding Percentage by School Type School Type Percentage

Public 72.9

Catholic 74.1

Other Private 60.2

Total 72.0

percent (in the urban/high Orshansky/small stratum) to a high of 84.2 percent (in the urban/low Orshansky/large stratum) (see Table 1). Table 4 shows that response rates for public and Catholic schools (72.9 and 74.1%) are highly significant when compared to response rates from other private schools (60.2%). The combined response rate for all schools was 72.0 percent. The resulting overall unequal weighting effect (UWE) for nT responding schools was 1.45, where

UWE n

~ (z^^r

Table 2 provides public school district-level eligibility and response rates for the three Orshansky levels. This table shows fairly equal eligibility rates across all levels with a slightly lower response rate (77.8%) for the lowest Orshansky level. The overall eligibility rate was 98.3% and the overall response rate was 80.3%. The overall non-response adjusted unequal weighting effect for school districts was 1.60.

5-4 Data Collection Experience Table 5 shows some weighted questionnaire findings from the teacher data collection activities. Approximately 64 percent of all public schools have a coordinator or a

166 Table 5. School Weighted Percentage of Teachers Responsible for Substance Use Prevention Education

School Type Percentage

Public 64.1

Catholic 42.2

Other Private 40.0

Total 56.3

Table 6. Weighted Percentage of Districts Responding Middle Schools with a Substance Use Prevention Education Coordinator

All Schools 67.4

Some Schools 9.2

No Schools 23.5

specific teacher who is primarily responsible for substance use prevention education at the school. This prevalence rate is significantly higher than that for Catholic and non-Catholic private schools (42.2 and 40.0 percent). Table 6 shows the weighted questionnaire findings for public school districts. Comparable to school findings, approximately 67 percent of public school districts indicated that a coordinator or a specific teacher was present in all schools within the district with the responsibility for substance use prevention activities in the school.

6

Conclusions

One of the most important sample design issues for this study relates to the identification of the sampling frame. When the SSUPPS study was designed, the data sources known were CCD, MDR, and QED. According to Hamann 5 , all three sources are adequate for public schools, but QED is the only source containing both public and private schools. This feature led to our use of QED. However, since this survey was implemented, NCES has made available on the NCES website (http://nces.ed.gov) a detailed data source for private schools known as the Private School Universe Survey (PSS). If the surveys were implemented today, we would recommend a combination of CCD and PSS for the sampling frame.

Appendix: List of Acronyms Used 1. SAMHSA - Substance Abuse and Mental Health Services Administration 2. SSUPPS - School-Based Substance Use Prevention Programs Study 3. SDFSCA - Safe and Drug-Free Schools and Communities Act

167 4. CCD - Common Core of Data 5. NCES - National Center for Education Statistics 6. MDR - Market Data Retrieval 7. QED - Quality Education Data 8. PSS - Private School Universe Survey

Acknowledgments The authors thank the supporter for this research: National Institute on Drug Abuse, Dr. Elizabeth Robertson (Project Officer), NIDA Grant 5R01 DA1149202. Thanks to members of the School-Based Substance Use Prevention Programs Study project team for comments concerning this document: Allison Burns, Susan T. Ennett, Matthew C. Farrelly, Ruby E. Johnson, Kurt Ribisl, Christopher L. Ringwalt, Luanne Rohrbach, Eva W. Silber, Ashley P. Simons-Rudolph, Judy M. Thorne, Amy A. Vincus, and Dana L. Wenter. In addition, the authors thank James R. Chromy for his suggestions and technical review, David Kellogg for his editorial review, and Brenda E. Gurley for her clerical contributions.

References 1. W.G. Cochran, Sampling Techniques, Third Edition (John Wiley & Sons Inc., 1977). 2. Dun &: Bradstreet Corporation, Market Data Retrieval Database(School Year 1994-1995). 3. L. Dusenbury and M. Falco, Journal of School Health 65, 420 (1995). 4. L.D. Eigen and D.W. Rowden, Making the Case for Prevention: A Discussion Paper on Preventing Alcohol, Tobacco, and Other Drug Problems (Department of Health and Human Services, Public Health Service, Washington, DC, 1993). 5. T.A. Hamann, Evaluating the Coverage of the U.S. National Center for Education Statistics Public Elementary/Secondary School Frame (Presented at the International Conference on Establishment Surveys - II. Governments Division, U.S. Census Bureau, 2000). 6. Jones, S.M., and J.R. Chromy, In Proceedings of the Section on Survey Research Methods (American Statistical Association, 1982). 7. L. Kann, J.L. Collins, B.C. Pateman, M.L. Small, J.G. Russ and L.T. Kolbe, Journal of School Health 65, 291 (1995). 8. L. Kish, Survey Sampling (John Wiley & Sons, New York, 1995). 9. Quality Education Data Inc., QED National Education Database: Data Users Guide, Version 4-6 Quality Education Data, Inc., Denver, CO, 1998).

168 10. C.L. Ringwalt, J.M. Greene, S.T. Ennett, R. Iachan, R.R. Clayton and C.G. Leukefeld, Past and Future Directions of the D.A.R.E. Program: An Evaluation Review (Prepared for the National Institute of Justice, 1994). 11. Substance Abuse and Mental Health Services Administration, Summary of Findings from the 1998 National Household Survey on Drug Abuse (Office of Applied Studies, U.S. Department of Health and Human Services, 1999).. 12. SAS Institute Inc., SAS Procedures Guide, Version 6, Third Edition (1990). 13. B.V. Shah, B.G. Barnwell and G.S. Bieler, SUDAAN Users Manual (Research Triangle Institute, Research Triangle Park, NC, 1996). 14. M.G. Sirken, Journal of the American Statistical Association 67, 224 (1972). 15. U.S. Department of Education, Public Elementary/Secondary Universe Survey (National Center for Education Statistics, School Year 1994-95).

O N E - S T E P ESTIMATION FOR THE PARTIALLY LINEAR P R O P O R T I O N A L H A Z A R D S MODEL XUEWEN LU Department of Mathematics and Statistics, University of Calgary, 2500 University Drive, Calgary, Alberta, T2N 1N4, Canada R.S. SINGH Department of Mathematics and Statistics, University of Guelph, Guelph, Ontario, NIG 2W1, Canada E-mail: [email protected] The proportional hazards regression model usually assumes that the covariate has a log-linear effect on the hazard function. In this paper, we consider a semiparametric survival model with flexible covariate effects. We assume the baseline hazard function can be parameterized, while the risk function associated with covariates is modeled in a semiparametric way. A "one-step algorithm" is used to estimate the nonparametric components and the parametric components by maximizing the local and the global likelihood, respectively. Using the local linear method, the estimates of the unknown parameters and the unknown covariate function are determined, and their asymptotic distributions are obtained.

1

Introduction

The Cox regression model (see Cox 7 ) is an important model in survival analysis, in which the conditional hazard of failure at time t given the covariate value z is modeled as h(z\t) =

h0(t)ex.p(zT/3).

Here, the baseline hazard ho(t) is nonparametric, while the dependency on the covariate z is parametric. The partial likelihood principle provides efficient estimates of (3 in the semiparametric model (see Andersen and Gill 1 ) . The Cox proportional hazards model has had profound impact on clinical trials for quantifying the effects of covariates on the survival time and for controlling confounding by means of a mathematical model which takes the outcome under consideration as the dependent variable, and includes both the postulated causal factor and confounding factors in the equation (see Fleming and Lin 13 and Elwood 1 0 ) . Despite the advantages of the Cox model, many authors have considered nonparametric and semiparametric modeling of covariate effects on the censored failure time. For example, Beran 3 , Dabrowska 8 , McKeague and

169

170

Utikal 2 3 and Nielsen and Linton 25 study the fully nonparametric hazard model where h(z, t) is unspecified. Fan, et al. n treat the multiplicative nonparametric model where h(z,t) = h0(t)X(z) in which neither h0(-) nor A(-) is parametrically specified. Sasieni 29 ' 30 examines a partially linear covariate effect in a model where h(z,t) = ho(t) exp{zf(3 + Xfa)}- Dabrowska 9 discusses a generalized Cox regression model with baseline intensity dependent on a covariate in which h(z,t) = h0(t, zi)exp{zj/?}. Hastie and Tibshirani 15,16 consider a fully nonparametric additive Cox model. Huang 17 studies a partly linear additive Cox model. Grambsch, Therneau and Fleming 14 and Fleming and Harrington 12 (Section 4.5, pages 163-168) propose to use smoothed martingale residuals to explore the functional form of the covariate effect in the Cox model. A survey of other regression models for censored survival data can be found in Andersen et al. 2. In this paper, we study the following semiparametric survival model : h(t\z, x) = h0(t; a) exp{zT/3 + \{x)},

(1)

where ho is a baseline hazard function with an unknown parameter vector a. (zT,x)T e R p x R is a covariate vector. The variable x is continuous with values in a compact set X £ R, and z is discrete or continuous with values in R p . 0 = (/3T, aT)T is an unknown q + p parameter vector, A() is an unknown smooth univariate function. In this model, the baseline hazard is modeled parametrically; the covariate effects are modeled semiparametrically. In many applications the shape of the baseline hazard is well understood: for example, the Gompertz-Makeham hazard is used extensively in insurance problems (Jordan 19 , page 21). A linear approximation to the logarithm of the baseline hazard is successful in a number of chronic diseases problems (Meshalkin and Kagan 2 4 ) . Also, the Weibull baseline is widely used in both biostatistical and reliability applications, e.g. Lawless 21 . The covariate effect, however, is rarely specified by a completely parameterized model. For instance, in a clinical trial study, when Z is a treatment covariate and X is a covariate describing characteristic of the patients, B can be interpreted as a measure of the treatment effect after adjusting for the effect of X. The effect of X can be modeled in a nonparametric way. Fleming and Lin 13 provide an overall review for the developments of survival analysis in clinical trials. Hence, this model allows flexible modeling of the covariate effect in many applications including clinical trials. We assume that the covariates are time-independent due to the fact that this type of data arises often in practice and the technical details involved in time-dependent covariate models are hard to tackle. Nielsen et al. 26 study model (1) without the linear part zTf3 and they assume the covariate x is time-dependent. As they mention in their paper,

171

the partial likelihood principle fails to estimate the parametric part in model (1). A generally applicable approach for this type of semiparametric model is that of profile likelihood, which is discussed extensively in Severini and Wong 31 and Bickel et al. 4 . When X(x) has a parametric form, for example, \(x) = xTj, it is well known that one can use maximum likelihood methods to analyze the model. Kalbfleisch and Prentice 20 and Lawless 21 give a detailed introduction to the use of maximum likelihood methods in the analysis of parametric regression models in the context of survival analysis. Most statistical software, for example, SAS and SPLUS, can analyze the Cox model, but they lack the ability to estimate the parameters in the baseline hazard function when the covariates are modeled in a nonparametric or semiparametric manner. That is, when the regression function A(a;) is not parameterized as a linear form or is an unknown regression function. This paper discusses how to estimate both the parametric and nonparametric parts in model (1). The paper is organized as follows. Section 2 of the paper introduces the likelihood for the survival model under right-censoring. Section 3 describes the local likelihood and local linear fit method. Section 4 proposes the onestep estimation method. Section 5 studies the asymptotic distributions for both parametric and nonparametric parts. Section 6 gives some conclusion remarks on the proposed method. Finally, the Appendix gives the technical proofs. 2

Likelihood Function for A Parametric Survival Model

Let f(t\z, x) denote the conditional density function of lifetime T given (Z,X) = (z,x), and let S(t\z,x) = P{T > t\(Z,X) = (z,x)} be its conditional survival function. The conditional distribution function of a random censoring variable C given {Z,X) = (z,x) is denoted by G(t\z,x). Then under independent and noninformative censoring (i.e., G(t\z, x) does not involve the unknown parameters), the conditional likelihood function is given by

L^UfWZuXjUsiYilZuXi), u

(2)

c

where Y[u a n d Ylc denote, respectively, the products involving the uncensored and the censored observations. Given that Ho(t;a) is the cumulative baseline hazard function and that \(x) is parameterized as A(a;) = A(a;;7), the conditional survivor function for T at (Z, X) = (z, x), under the survival model (1), is of the form S(t\z, x; a, (3,7) = e x p [ - # 0 ( i ; a) exp{zT/3 + X(x; 7)}].

172

The hazard function and cumulative hazard functions are given by h(t\z,x;a,p,j) = h0(t;a)exp{zTP + X(x;j)} and H(t\z,x;
L{0,l) =

HihiYilZuXi^SiYilZuXi), 1

remembering that 6 = (/?T, <jT)T. Under the survival model (1), we have the log-likelihood for the sample

i„(«.7) = S{«iiog/(yi|zilxo + (i-«i)iogS(yi|zi,xi)} i

= ] > > { l o g h0(Yi; a) + {Z?P + A(X i; 7))} i

+ exp{Z?P +

\(Xi;1)}logSo(Yi;cj))

= ^ [ « i { l o g h0(Yi; a) + {Zf (3 + A(X i; 7))} i

-exp{Z?P

+ \(Xi;i)}Ho{Yi;iT)],

(3)

using Ho(t;a) = — logSo(t; a). Maximization of (3) leads to the maximum likelihood estimators of a, (3 and 7. 3

Local Likelihood

Suppose that the form of X(x) is not specified, and that the first order derivative of X(x) at the point x exists. If we let ao — X(x), 01 = X'(x), then, by Taylor's expansion, X(X)aa0+a1{X-x), for X in a neighborhood of x. Let b be the bandwidth parameter that controls the size of the local neighborhood and let W be a kernel function. Using this local model, one would find ao, a^ and 6 = (/3T,aT)T to maximize the local (log) likelihood n

ln(a0,a1,(3,a)=Y,k{
+ ao +

a1{Xi-x),(Yi,6i)}Wb{Xi-x),(4)

173

where h{a, Zj(3 + a0 + ax{Xi - x), (Y{, Si)} = &[logh0{Yi; a) + {ZT/3 + a0 + - x))] - Ho{Yi;a)eXp\ZT(3 + (Zf'/? + a0 + ax{Xi - x))}, Wb{t) = ai{Xi b-lW(t/b). But the estimates of the true parameters 9Q = (f3j, &£)T from this local likelihood are not efficient since we use only a fraction of all data points. Lu et al. 22 consider model (1) using the generalized profile likelihood method with local constant fit and obtained an efficient estimate of (/?o, 0"o) which achieves the semiparametric information bound. But that approach requires a fully iterated algorithm and is not computationally convenient. In this paper, we provide the one-step algorithm with local linear fit. Application of the local linear fit enables us to reduce the bias of the estimator for the nonparametric component and to avoid boundary effects. In some important applications, we show that one-step algorithm achieves the same efficiency. 4

The One Step Estimation Method

Under model (1), the primary interest is to estimate the true parameters
Y.lifaZj'P

+ XiXi-MAYuSi)}.

(5)

i=l

Step 3: Fix 9 at its estimated value from Step 2. The final estimate of A0(-) is A(a;; b, 9) = a 0 where (So, &o) maximizes n

Y^ k{°, ZU + ao + ai(Xi - x), (Yi,6i)}Wb(Xi i=l

- x).

(6)

174

At this final step we take b to be an estimate of the bandwidth that is optimal for estimation of Ao(-) when 90 is known. To compare the one-step algorithm with the generalized profile likelihood method or the fully iterated algorithm, we also give the fully iterated algorithm as follows: Step 0: Fit a parametric linear model to obtain the initial estimates 9. Step 1: Find A(a:; b, a, /3) by maximizing the local likelihood (4). We take b to be an estimate of the bandwidth that is optimal for estimation of 9Q. Step 2: Update 9 by maximizing n

J2 hW, ZTf3 + \(x; b, a, 0), (Yh Si)}.

(7)

Step 3: Fix 6 at its estimated value from Step 2. The final estimate of Ao(-) is \(x; b, 9) = a$ where (&o, oi) maximizes (6). At this final step we take b to be an estimate of the bandwidth that is optimal for estimation of Ao() when 6Q is known. Instead of maximizing (7), an alternative approach is to update the current 6 by maximizing n

Y.idczTp

+

Xfab&JiiYiji)}.

i=l

We conjecture that the estimators resulting from this type of updating are equivalent to those obtained by maximizing (7); this would be interesting to verify. 5 5.1

Distribution Theory One-Step Estimator of the Nonparametric Part

In this subsection, we investigate properties of the estimators of the nonparametric part Ao(-) of (1) under two cases: (1) the one-step approach when 9Q is estimated locally as in (4); and (2) the final estimate as in (6) when 9Q is estimated globally. In case (1), we are using only 0(nh) data points to estimate both 90 and Ao(-). In case (2), #o is estimated at parametric rates, and thus AQ(-) can be estimated asymptotically as well as if #o were known.

175 We consider case (1) first. Let /(•) be the density of X. Define

X = x

£(*)

mi = Ao(Xi) + Z?P;

(

{Si - HQ(Yi; (To) exp(mi)}Ui {Si - H0(Yi; do) exp(mj)}Zi 6ih'0{Yi;CT0)- H'0(Yi;CT0)exp(m;) y

i>(:r) = first diagonal element of the matrix S _1 (a;); ho(t;
and ft,'0(t;(T) — dho(t;cr)/d<J.

Theorem 5.1. Under condition A given in the appendix, as n —> oo, b —> 0 and nb —> oo and n&5 is bounded, for the the maximizer of the local likelihood (4), we have (nb) 1/2

\(x) - A0(x)

P~Po a -

«2

-^A^E-1^

|A" = a;

(T0

N 0, Hence (^) 1 / 2 {A(x) - A0(x) - ^A 0 '(:r)& 2 } ^ N(Q, ^o-£?v{x)).

m

5.2

(8)

Fully Iterated Estimator of the Nonparmetric Part

One-step estimation of the parameters
176

Theorem 5.2.

(nb)^{\(x)~X0(x)-^b2}^N

0,

f(x)E(S\X = x)

The asymptotic variance in this result coincides with the univariate result given in Nielsen et al. 2 6 and Lu et al. 22 when covariates are time-independent, but the bias here is different, because we use the local linear fit. This result suggests bandwidth estimators in the spirit of that of Ruppert, Sheather and Wand 2 8 . Following the discussions made in Carroll et al. 5 ' 6 , for example, consider estimation of Ao(-) at the final step, for a given function w(-) with compact support, minimizing the asymptotic weighted mean squared error with weight f{-)w(-) yields the optimal global bandwidth bopt = C(W)n

/ v*(x)w(x) dx "J f\%(x)2f{x)w{x)dx

-1/5 | .

where C(W) = ( ^ 0 K ^ 2 ) 1 / 5 . For the one-step algorithm in Step 1, a relatively ad hoc choice of b is bopt x n 1 / 5 x n " 1 / 3 = bopt x n _ 2 / 1 5 . On the other hand, for the fully iterated algorithm (7), we have shown that ordinary bandwidth rates are permissible (Lu et al., 22). Severini and Staniswallis 3 2 , Hunsberger 18 , Severini and Wong 31 show the same thing for generalized partially linear model. 5.3

One-Step Estimator for the Parametric Part

We now study the global estimators for the parametric components

(CTOJPO)-

Theorem 5.3. Let J3 and a be the one-step estimates which maximizes the likelihood (5). Under conditions A given in the appendix, as n —• oo, nb4 —» 0 and nb2/ log(l/6) —> oo,

nW^-M^o.ir^ir1), \ a - aa J where B = Ex=B

+

E(6VVT), E{1(X)1T(X)ejE-1(X)eJ},

-y(x) = E(6V\X = x), V = {ZT, h'Q(Y;cr0)T)T, the first position.

ex is the unit vector with 1 in

177

Therefore, one iteration leads already to a root-n consistent estimator of 0o = (/3 0 Vo T ) T We have derived the results using local constant fit for the fully iterated estimator defined by (7) (see Theorem 5.2, Lu et al., 2 2 ) . The results also hold for the local linear fit in this paper. For the sake of completeness, we cite the result as follows: Theorem 5.4. Let 6 = (/3T, aT)T be the fully iterated estimates which maximizes the likelihood (7). Under Condition I, S and A given in Lu et al. 2 2 , assume ig0 is positive definite and finite, then

where ig0 is the marginal Fisher information matrix for 6Q given by dl,„ ieo=E0^—(60,\0)

,

v

+

81

x

—(e0,\0)(v*)

®2

v* = X'go is the least favorable direction. ig0 can be consistently estimated by ig0 has the following simple expression: ie0 = E0 j ^ | g r ( t , Z, X)h0(t; a0) exp[Z T /3 0 + X0(X)]I(t)

dt

-E0J^^(t,Z,X)dN(t) =

Eo[6{d-^dJ^(Y,Z,X)}}

where n8{t,z,x) = log[/i o (i;0-)exp{z T /3 + Xg(x)}}, Xe(x) = X0(x) + rBo(x) rg(x) and rg{x) = \ogE{exp(ZT(3)Ho(Y;cr)\X = x}. We have proved ig0 is the semiparametric information bound. R e m a r k 5.1. The fully iterated estimator is uniformly as or more efficient than the one-step estimator. However, when X and Z are independent and 6 = 1 (there is no censoring), the one-step estimator is as efficient as the fully iterated estimator. Therefore, iteration is not necessary when X and Z are weakly correlated and the censoring is not heavy. In fact, if we set E(Z) = 0 without loss of generality, then, when X and Z are independent and 6 = 1, we have E(SZ\X) = E(Z) = 0, and the asymptotic variance for the one-step estimator of /? equal to {E(ZZT)}~1, which is also the asymptotic variance for the fully iterated estimator given in (7).

178

6

Conclusions

For the generalized partially linear single-index models, the asymptotic distribution theory for the one-step and the fully iterated estimator, based on the local linear fit, has been established by Carroll et al. 5 . In this paper, we formulate our approach in the frame work of proportional hazards regression models. We can draw some similar conclusions from our study: (i) estimation of the nonparametric part Ao(-) of model (1) can be done just as well as if the parameters (<70, /?o) were known; (ii) the parametric parts can be estimated at the usual parametric rate of convergence; (iii) the fully iterative estimator can have a smaller variance-covariance matrix than the one-step estimator, but for the case without censoring and with weakly correlated X and Z both estimators share the same variancecovariance matrix; (iv) estimation of the nonparametric part in the model has an effect on the distribution of the estimates for the parametric parts; (v) best estimation of the parametric parts requires undersmoothing of the nonparametric part. Our model (1) is restricted on univariate in the nonparametric part. When there are several covariates in the nonparametric part, dimensionality reducing strategies such as additive and index models can be used. For instance, one could fit the following partially linear single-index model: h{(t; z, x) = h0(t, a) exp{zT(3 + A(x T 7 )}, where A(-) is of unknown form and, a, /3 and 7 are unknown parameters. Further development of this issue needs to be investigated. Appendix: Technical Proofs The following conditions will be needed: Condition A: (i). 6Q = (
179

(iii). There exists £ > 0 such that E[{H0(Y; <70) exp(Z r A,)} 2 +«|X],

E{\H0(Y; a0)

exp(ZTf30)Z\2+^\X}

and E{\h'0(Y;ao)\2+t\X},

E{\H'0(Y;a0)exp(ZTPo)\2+i\X}

are finite and continuous at the point X log h0(Y;a0).

= x, where ho(Y;(To) =

(iv). The functions E{H0(Y; a0) exp{ZT/30)\X}, E{H0(Y; <x0) exp{ZTPo)\X}, E{H'0(Y; a 0 ) exp(Z T po)Z\X), E{H0(Y; a0) exp(ZT/30)ZZT\X}, T E{H'Q'(Y;a0)exp(Z p0)\X}, E{fh'0{Y;a0)\X} and E{6h%(Y;a0)\X} are all continuous at the point X = x. The function E(5\X) is positive and has a continuous 2nd derivative around the point x . (v). There exists a function M(y), with E{M(Y)} 3

d

daAdakda{

< oo, such that

3

h0{y,(j)

< M(y),

d

dcrjdffkdcTi

•H0{y;c)

< M(y),

for all y, and for all cr in a neighborhood of OQ. (vi). The kernel function W > 0 is a bounded density supported on [—1,1] symmetric about zero. (vii). The function XQ(X) has a continuous 2nd derivative around the point (viii). The density /(•) of X has a continuous 2nd derivative around the point x and f(x) > 0. Let go — ho(Y;ao),

define

S0(x) = E < {

0 K2 0 0 X Z 0 ZZT Zgl \<7o 0 g0ZT gf2 )

and

Si(x)

=E<

( v0 0 v0Z V go

0 VQZT v0go \ v2 0 vo0 0 vaZZT v^Zgl 0 <70ZT ^ o ^ 2 /

X =, > .

180

Proof of Theorem 5.1: Let cn = (n/i)" 1 / 2 , 1 \ / c-^a-Xoix)} X! = \(Xi - x)/b ) , /3* = I c~\{& - A'0(o:)}

, Ui=(

l

)

and let a* = cn {o - o-o), Denote A{ = A^o;) = Ao(a;) + Zf(30 + X'0(x)(Xi — x). If (ao, a\,6)T mizes (4) then 9* maximizes

maxi-

5 3 Z;(cra<x* + (T0, cnp*TX* + Xi, (Yit 6i))Wb(Xi - x) with respect to 0*. Consider the normalized function n

/„(r) = ^53[zi(c„(r*+fT0,cni3*Tx;+Ai,(yi,<5i))-/i((To,Ai,(yi,^))]wb(xi-x), t=i

which is maximized by 6*. ln(9*) is a concave function, we can apply the Convexity Lemma (see Pollard 27 ) to it. By a Taylor expansion of the function h{-, -, {Yi, Si)) we obtain that

in(9*) = wis* - ~e*TAne*{i +

0p(i)},

0)

where n Wn = hcn ^

/{6iHo{Yi;a0)exp(Xi)}Ui {6i - H0(Yi; a0) exp(Ai)}^

i=i \6ih'0(Yi;
_

] Wb(Xi - x)

- if 0 (y i; CTo)exp(Ai)

and _iHo(n •>0)ZiVj i = 1

\ e ^ ^ ( V ,

c * i H ( | ( y i ; » o ) 2 i 2 j ' oXi ZiHla(Yi;"Si)r \°0)ZT -6ih'^Yi;<,0) +

) Wb(Xi

~ x).

Hl^Yi;^0)ex

It is shown that An = f(x)S0(x)

+ o p (l) = A + o p (l).

Therefore, by (9), ln(G*) = WZe*-^e*TA6*

+ op(l).

(10)

181

By applying the Convexity Lemma, we obtain that 0*=A-1Wn

+ op(l).

Hence, the asymptotic normality of 9* will follow from that of Wn. Since Wn is a sum of i.i.d. random vectors, we only need to compute the first two moments and check Liapounov's conditions. It can be shown that E{Wn) = Vri{f(x)^\'0\x)b2 Var(Wn)=f(x)S1(x)

+ o(b2)}, + o(l).

Therefore, we have Wn-EWn^N{0,f(x)S1(x)}. Proof of Theorem 5.3: We need two lemmas, Lemma A.l and Lemma A.2, given by Carroll et al. 5 in order to prove Theorem 5.2. By Lemma A.2, each element in An converges uniformly to its corresponding element in A. Hence, expression (9) holds uniformly in x G X. By the Convexity Lemma, it also holds uniformly in 6* G C and x G X for any compact set C. Lemma A.l then yields sup\6*{x) - A~lWn{x)\ xex

^ Q,

(11)

where 9*(x) and Wn(x) are defined in the proof of Theorem 5.1, both depend on x. By considering the first element of the vectors in (11), we have 1

— V WiWb(Xi - x) = sup X(x) - Ao(ar) - -nf(x) ^ xex where W; is the first element of the vector E-1^)

op(cn),

f{6i-H0(Yi;a0)exp(Xi)}Ui {6i-Ho(Yi;
By a result of Carroll et al. 5 , the following stronger result holds: L - £ WiWb(Xi - x] sup X(x) - \0(x) - — nf(x) t = i xex Let §! = nll2C$ - /?o), k = n^2(a rhi = \(Xi) + Zfpo,

0P{b2cn

+

c2n\og^2(l/b)}.

- a0), and 9 = (Of, 6%)T. Denote by

and

rm = A0 {X{) + Zj (30 •

182

Then, 9 maximizes n

UO) = ^ [ ^ ( " " 1 / 2 ^ i + °o, n-l'2Zf92

+ rhu (Yu S{)) - h(a0, mu (Yu Si))],

By Taylor's expansion, we have ln(9) = n-1'2

£

(Yi> 5 ^ ~ \tTBj,

tffc,

(12)

yhere <*, ( V . i{miAxl,oi}}

{6i-H0(Yi;a0)exp(mi)}Zi - F0(yi;a0)exp(mi)

K \ \ - {

- \s.^Yi;
where b\x = exp(m; + n 1^2Zi9ii)H0(Yi;ao + n 1/292i)ZiZ?, b[2 = exp(rhi + n-^ZiO^ZiH'oiYi; a0 + n~1/292i)T, b21 = exp(m i + n-1^Zi91i)H'0(Yi; a0 + n~V292i)Zj and b\2 = - & / # ( * ; ;
Bn

E{SZh'0(Y;a0)T} E{S(h'0(Y;a0))®2})

\ +

° ^ 1 } = ~B + °p{1>-

As for the first term in (12), we have n

n~1/2 £

n

i{mi, (Yh Si)) = n-1'2 £

1=1

^(rm,

(YuSi))

1=1 n

+

n-ll2Y,^J{mi,{Yi,Si)){\{Xi)-\0(Xi)} t=i 1 2

+ Op(n / ||A-A 0 ||^ o ), where Mrnu (Yi, Si)) = ^(rm, dmC

(Yi, Si)) = - ( Hf^ f * ) . \ H0(Yi;
183 The second term in the above expression can be expressed as n

n

n -3/2 £

^

^

{Yu

1 - Xi) ( 5 i ) ) / ( x i ) - Y, WjWbiXj

t=l

1=1 l 2 2

+0P{n / c Define z^- = v(Xj,Zj,

(

l o g ^ l / f c ) } = Tnl + 0P{n^cl

n

log^l/fc)}.

(Yj,6j)) as the first element of

Sj - HQ(YJ;
Using the definition of Xj(Xi), therefore n

Zj

dM(t).

\h0(t;a0)J

we obtain A, — rrij = 0((Xj

— Xi)2)

and

n

Tnl = n- 3 / 2 5 3 ^ 2 Vf K , (YuSiWiXiy^MXj

- Xi) + 0P{nl'2h2)

i=lj=l

By a similar procedure used in Carroll et al. 5 , it can be shown that T„ 2 - T„ 3 £ 0,

(13)

where

j=i

with

IV i?o(yi;cro)exp(??2i) y 1

j

It is seen that

Hrni^YiA))

= j
dMi{t)

Combining (12)-(13) we obtain that n 1 2

Zn(0) - n " / ^ J2 tyXi, (Y, Si), Zi) - -6TB9 + oP(l), t=i

184

where

n(xu (Yh Si), Zi) = Mm, W, st)) - 7(*iH By the Convexity Lemma we have n

from which it follows that 6^

N(Q,B~llliB-1).

This establishes the result in Theorem 5.3. Acknowledgments Research of R.S. Singh is supported in part by the National Science and Engineering Research Council Grant # A4631. References 1. P.K. Andersen and R.D. Gill, Ann. Statist. 10, 1100 (1982). 2. P.K. Andersen, O. Borgan, R.D. Gill and N. Keiding, Statistical Models Based on Counting Processes (Springer-Verlag, New York, 1993). 3. R.J. Beran, Nonparametric regression with randomly censored survival data (Technical report, Univ. California, Berkeley, 1981). 4. P.J. Bickel, C.A.J. Klaassen, Y. Ritov and J.A. Wellner, Efficient and Adaptive Inference in Semiparametric Models (Johns Hopkins Univ. Press, 1993). 5. R.J. Carroll, J. Fan, I. Gijbels, and M.P. Wand, Discussion Paper #9506 (Institute of Statistics, Catholic University of Louvain, Louvain-la-Neuve, Belgium, 1995). 6. R.J. Carroll, J. Fan, I. Gijbels, and M.P. Wand, J. Amer. Statist. Assoc. 92, 477 (1997). 7. D.R. Cox, J. Royal Statist. Soc. 34, 187 (1972). 8. D.M. Dabrowska, Scand. J. Statist. 14, 181 (1987). 9. D.M. Dabrowska, Ann. Statist. 25, 1510 (1997). 10. J.M. Elwood, Critical appraisal of epidemiological studies and clinical trials, Second Edition (Oxford University Press, London, 1998). 11. J. Fan, I. Gijbels and M. King, Ann. Statist. 25, 1661 (1997). 12. T.R. Fleming and D.R. Harrington, Counting Processes and Survival Analysis (John Wiley & Sons, New York, 1991).

185

13. T.R. Fleming and D.Y. Lin, Biometrics 56, 971 (2000). 14. P.M. Grambsch, T.M. Therneau and T.R. Fleming, Biometrika 77, 147 (1990). 15. T. Hastie and R. Tibshirani, Generalized Additive Models (Chapman and Hall, London, 1990). 16. T. Hastie and R. Tibshirani, Biometrics 46, 1005 (1990). 17. J. Huang, Ann. Statist. 27, 1536 (1999). 18. S. Hunsberger, J. Amer. Statist. Assoc. 89, 1354 (1994). 19. C.W. Jordan, Life Contingencies (The Society of Actuaries, Chicago, 1975). 20. J.D. Kalbfleisch and R.L. Prentice, The Statistical Analysis of Failure Time Data (John Wiley and Sons, New York, 1980). 21. J.F. Lawless, Statistical Models and Methods for Lifetime Data (John Wiley & Sons, New York,1982). 22. X. Lu, R.S. Singh and A.F. Desmond, Journal of Statistical Planning and Inference 98, 119 (2001). 23. I.W. McKeague and K.J. Utikal, Ann. Statist. 18, 1172 (1991). 24. L.D. Meshalkin and A.R. Kagan, Discussion of "Regression models and life tables" by D.R. Cox. J. Roy. Statist. Soc. Ser. B 34, 213 (1972). 25. J.P. Nielsen, O.B. Linton, Ann. Statist. 23, 1735 (1995). 26. J.P. Nielsen, O.B. Linton and P.J. Bickel, Ann. Statist. 26, 215 (1998). 27. D. Pollard, Econometric Theory 7, 186 (1991). 28. D. Ruppert, S.J. Sheather and M.P. Wand, J. Amer. Statist. Assoc. 90, 1257 (1995). 29. P. Sasieni, Scand. J. Statist. 19, 215 (1992). 30. P. Sasieni, J. Roy. Statist. Soc. Ser. B 54, 617 (1992). 31. T.A. Severini and W.H. Wong, Ann. Statist. 20, 1768 (1992). 32. T.A. Severini and J.G. Staniswalis, J. Amer. Statist. Assoc. 89, 501 (1994).

A N E S T E D FRAILTY SURVIVAL MODEL FOR R E C U R R E N T EVENTS RENJUN MA Department of Mathematics and Statistics, and Canadian Research Institute for Social Policy, University of New Brunswick, Fredericton, Canada, E3B 5A3 E-mail: [email protected] J. DOUGLAS WILLMS Canadian Research Institute for Social Policy, University of New Fredericton, Canada, E3B 5A3 E-mail: [email protected]

Environmental

Brunswick,

RICHARD T. BURNETT Health Directorate, Health Canada, Ottawa, Canada, KlA E-mail:[email protected]

0L2

We consider a nested frailty Cox proportional hazards model for recurrent events. We adopt a Poisson modelling approach and the principal results depend only on the first and second moments of the unobserved frailties. The use of the proposed methods is illustrated through the reanalysis of chronic granulotomous disease data previously reported by Fleming and Harrington 5 .

1

Introduction

Recurrent events often arise naturally in practice. Examples of such events include disease onsets and emergency room visits in medical studies, process stoppages in reliability studies, as well as criminal offences and unemployment in social studies. Such recurrent events are often clustered by subjects and their residential areas or other groupings. Until recently, the recurrent events are usually modelled by incorporating a single level of subject frailties, or random effects, into the survival models, while ignoring other level of clustering effects (Yau 1 6 ) . Furthermore, the existing approaches to frailty survival models generally rely on specific distributional assumption of frailties such as gamma, inverse Gaussian and log-normal (Sastry 13 , Yau 1 6 ) . In this paper, we consider a nested frailty Cox proportional hazards model for recurrent events where the nested frailties are specified only up to the first and second moments. The treatment of ties and stratification is explicitly incorporated in our approach in the same way as in the standard Cox model. Our estimation procedure is based on a characterisation of this frailty Cox model as an auxiliary random effects Poisson regression model (Ma et al. u ) .

186

187

This approach gives optimal and consistent parameter estimators based on the orthodox best linear unbiased predictor approach to the auxiliary random effects Poisson models. For a single level of gamma frailty, given the frailty parameter, our approach coincides with the hierarchical likelihood approach (Ha et al. 6 ) and the EM algorithm approach (Klein 10 , Nielsen et al. 12 and Andersen et al. 2 ) . We introduce the nested frailty Cox proportional hazards model and its auxiliary random effects Poisson models in sections 2 and 3, respectively. In section 4, we discuss the estimation of the nested frailty Cox models, using the orthodox best linear unbiased predictor approach to the auxiliary random effects Poisson models. An application of our approach to the chronic granulotomous disease data is illustrated in section 5. Some alternative frailty models are discussed in section 6. Some comparisons of our approach with others in the literature are discussed in section 7. 2

Nested Frailty Cox Model

In this section, we consider a Cox model with two levels of nested Suppose that the cohort of interest is composed of m independent indexed by i. Within the ith cluster, there are Ji correlated subjects by (i,j). Specifically, we assume that the cluster-level frailties Ui,..., positive, independent and identically distributed with E(Ui) = 1 and var(Ui) -a2,

frailties. clusters indexed Um are

i - l,...,m.

(1)

We also assume that there is a subject frailty Uij associated with the subject j from the ith cluster, j — 1,..., J;; i = 1 , . . . , m. We further assume that, given the cluster-level frailties U* = u* = (u\,... ,um), the subject-level frailties Wij},j = 1, ••., Ji',i = l , . . . , m are positive and conditionally independent, and that the conditional distribution of Uij, given {/* = u*, depends on Ui only, with E(Uij\U* = u«) = Ui and var(Uij\U* = «») = u2Ui,

(2)

j = l,...,Ji and i = l , . . . , m . Furthermore, assume that there are n^ recurrent events within the subject (i,j) where the A;th recurrent event time of the jthe subject in the ith cluster is given by y

_ J time to infection/censoring (^ time to infection/censoring

first episode (k = 1) since latest event (k > 1)

Let the hazard function for the fcth recurrent event time of the jth subject in the ith cluster at time t be denoted by Xijk{t). Given both the cluster

188

and subject level frailties U = u, we assume that the hazard functions are conditionally independent, with A«*(t) = ^k)(t)uij

exp{pTxijk(t)},

(3)

where k = 1,2,..., a with a being the maximum number of observed recurrences of the event, where AQ (t) denotes the unspecified baseline hazard function (Cox and Oakes 4 ) and Xijk{t) denotes the time-dependent covariate vector for the kth recurrent event time of subject i, j at time t. The components of this time-dependent covariate vector Xijk (t) will be approximated by piecewise constants described in the next section. The distribution of frailties is assumed not to depend on the regression parameter /?. Clearly, the recurrent event times, either observed or censored, within the same cluster are correlated. An implication of this model is that the subject riij is not at risk of (k + l)th event until this subject has experienced the fcth event. A Cox proportional hazards model with a single level of frailties is obtained as a special case of the Cox model with two levels of frailties by setting v2 — 0 and Ji — 1 for all i. Our model only requires the specification of the first two moments of frailties since it is generally impractical to assume that the random mechanism by which the unobserved frailties were generated is entirely known. On the other hand, our assumptions (1) and (2) on random effects do cover a wide range of frailty distributions including gamma, inverse Gaussian and log-normal. 3

Auxiliary Random Effects Poisson Models

Let Sk be the set such that (i, j) € Sk if and only if the subject (i,j) has experienced his (fc-l)th recurrent event time. Let TH < . . . < Tkgh denote the distinct fcth recurrent event times in Sk, with rrikh indicating the multiplicity of fcth recurrent events occurring at time 7>/, (k — 1 , . . . , a). The risk set at time Tkh is a subset of Sk such that TZ{Tkh) — {(i,j) '• Ujk > Tkh}, where Ujk is the feth observed recurrent event time for the jth subject in the ith cluster. In addition, for any (i,j) € 5^, let Yy/b,/, be 1 if the fcth recurrent event occurs for subject (i,j) at time Tkh and 0 otherwise. Let Y and U denote the vectors of the Yijk,h and the random effects (frailties) Uij, respectively. Given the random effects U = u, Peto's version of the conditional partial likelihood (Cox & Oakes, 1984, p. 103) is / (B-Y\u) - 1 7 W

ft

' ^ MM

n ^ ^ ^ ^ ^ ^ i e x p ^ ^ ^ ) ) } ^ , , ^ ! )

iZiiJWr^eM^iATkh))}^

' (4)

189

where the time-dependent covariate Xijk(t) is approximated by a piecewise constant function Xijk(t) - xijk(Tkh) t e [rk{h-i), Tkh] for h = 1 , . . . , qk with Tfco = 0. The value of Xijk{Tkh) can be taken as any reasonable quantity such as the median of Xijk(t) over the interval (^^-1),^^]We now define an auxiliary random effects Poisson regression model. Assume that the components of Y are conditionally independent, given random effects U = u, with Yijk,h ~ Poisson{itij exp{akh + prxijk(Tkh))}

{i,j) & Tl{Tkh).

(5)

This auxiliary random effects Poisson model extends that of Whitehead (1980) for standard Cox models to frailty Cox models. Given the random effects, the conditional likelihood for the random effects Poisson model is a

qk

£li M

eX

P{E(i,i)eK(r fch ) UH ex P( a fch + PTXijk(Tkh))}

'

(6)

It has been shown by Ma et al. 1X that

t i Uii

mkh

-

J

where the term in parentheses on the right-hand side does not depend on the parameters of interest. This demonstrates that the maximum joint Poisson likelihood estimators for the regression parameter vector /3 from (6) are the maximum joint partial likelihood estimators for the regression parameter vector P from (4). We may therefore make inferences on the frailty Cox models by fitting random effects Poisson models. In the remainder of this paper, we focus on the nested frailty Cox proportional hazards models specified by (1), (2) and (3) by way of fitting the auxiliary nested random effects Poisson models specified by (1), (2) and (5). 4 4-1

Orthodox Best Linear Unbiased Predictor Approach Prediction of random effects

We predict the random effects by the orthodox best linear unbiased predictor of U given Y. If we let U and Y be random vectors with finite second moments, the orthodox best linear unbiased predictor of U given Y is

u = E(U) + Cov(t/,y)(Cov(r))-1 {Y - E(Y)} .

190

This is the linear unbiased predictor of U given Y which minimises the mean squared distance between the random effects U and their predictor within the class of linear functions of Y. The cluster random effects predictor can be expressed as fT _ i

1 + CT

-1 _i

Sfc=l Eft=l S(i,j)€TC(rth) 2 V^

a

V^^

fc

w Y

ij ijk,h

V"*

'

^ '

where (i, j) runs over the risk set TZ(Tkh) for any given i. Here, Vijk,h = exp (akh +

PTxijk)

= exp {(aT,0T)xijk,h}

,

and, for fixed (i, j),

(

a

1+

qk

^ 5353 k=i h=i

53

Vijk,h

(i,j)en(Tkh)

The right-hand side of the equation (7) clearly involves unknown quantities 7, a2 and v2. These quantities together with random effects predictors will be iteratively estimated. At each iteration, the Ui will be updated by evaluating those unknown quantities on the right-hand sides at their current values. The detailed discussion on the iterative algorithm will be presented in section 4.4. The unknown quantities involved in the following equations will be understood similarly. The subject random effects predictors are a

Qk

2

Uij =WijUi + V WiJY^'%2 k=l h=l

4-2

53

Y

ijk,h-

(8)

(i,j)e1Z(rkh)

Estimation of regression parameters

Consider first estimation of the regression parameters in the case of known dispersion parameters. Estimation of unknown dispersion parameters will be discussed in section 4.3. Differentiating the joint log-likelihood of the auxiliary model for the data and random effects yields the joint score function. Replacing the random effects with their predictors, we have an unbiased estimating function for the regression parameters 7 = ( a T , / ? T ) T : a

9k

^(7) = 53 5 3 fc=l h = l m

53 (i,j)€K(Tkh)

X

W<* {YHk,h - Uii{l)»ijk,h{l)\

191

This estimating function 1/1(7) is then a vector function of the same dimension as that of 7. The second equality is simply a rearrangement of terms under summation by clusters where a

«k

*/>i(l) = E E

E

Xi k h Yi k h

i> \ i,

- Uij(l)Hijk,h(l)}

corresponds to the ith cluster. For each i, it is clear from (7) and (8) that the random effect predictor Uij (7) is unbiased and involves only the induced observations, yijk,h, within the ith cluster, and so is the estimating function ipi(y). Therefore the unbiased estimating functions ipi{'y),---,il>m('y) are independent. The solutions of the estimating equation ^2^L1 V'JIT) = 0 provide estimators of the regression parameters. Noting that the dimension of parameter 7 increases with the number of clusters, the standard estimating equations theory based on the fixed number of parameters would apply if the dimension of parameter 7 is bounded. Otherwise, it can be shown that, under some regularity conditions, the component-wise asymptotically normality of parameter estimator, 7, remains valid if the dimension of parameter 7 grows slowly (He and Shao 7 ) . Specifically we have, for any scalar vector b of dimension of 7 such that bTb = 1, 6 T 7 is asymptotically normal with asymptotic mean bTj and asymptotic variance given by & T 5(7) - 1 V(-y)S('y)~rb. Here, the sensitivity matrix 5(7) and the variability matrix V(y) are given by

*r)=i><7)=i:s,{^}, i=l

i=l m

vh) = E i=l

K

!

J

m

v

^ = E ^ {^(7)^(7)} • t=l

It has been verified that S(^) — — V{p() for the nested random effects Poisson model (Ma et al. u ) ; therefore the asymptotic variance of bTj is simply - 6 T 5 ( 7 ) _ 1 6 . If the dimension of parameter 7 is bounded, the regression parameter estimator, 7, itself can be shown to be asymptotically normal with asymptotic mean 7 and asymptotic variance given by — -S'(7)_1 under mild regularity conditions (Artes and J0rgensen 3 ) . The estimating function Y^iLi V'iM c a n easily be shown to be optimal in the sense that it attains the minimum asymptotic covariance for the estimator bTj among a certain class of linear functions of Y. This estimating equation Y^T=i "4>i{l) — 0 c a n ' De solved by the Newton scoring algorithm introduced by J0rgensen et al. 9 : 7*=7-^1(7M7),

192

where 7* denotes the updated value for 7. Here negative sensitivity matrix plays a role similar to that of the Fisher information matrix in the Fisher scoring algorithm. In fact, the negative sensitivity matrix here corresponds to the so-called Godambe information matrix. The exact expression of the sensitivity matrix is given by m

«=1 j = l

i-\

qk

a —

Ji

m

^,2-f

Z-(

Viik,hXiik,h{Xijk,h)

,

(9)

fc=i />=i (i,j)en(Thh) where

(

a

qk

5Z ] C

\

w

ijtJ'ijk,hXijk,h

5Z

,

*=1 h=l (ij)€7e(rfch)

/«J

=

I 2 J Z_/

X/

(10)

/

V>ijk,hXijk,h J •

\fc=i h=i (i,j)en{Tkh)

J

Here, the index (i,j) in (10) runs over the risk set IZfah) for fixed i, and (i, j) runs freely over the risk set Tl(Tkh) in the last term of (9). In addition, c(i) denotes the mean squared distances between the random effect Ui and its predictor, specifically a = E(Ui - Uif a2 1+

a

2_)fc=l 2J/i=l z2(i,j)<EK(Tkh) wijixiik,h

where (i,j) runs over the risk set TZ(rkh) for fixed i. An analog of Wald's test is available for testing the hypothesis Ho : /3(!) = 0, where f}(i) is a subvector of /3. The test statistic is ^ = %{J11(7)}"1^(i), where J11 (7) is the block of the asymptotic covariance matrix of 7 corresponding to /?(!). Asymptotically, W follows a x2(fc) distribution, where k is the size of the subvector /?(i).

193

4-3

Estimation of dispersion parameters

We now suppose that the dispersion parameters are unknown. By analogy with generalized linear models, we adopt the following adjusted Pearson estimator for the dispersion parameter a2: m

i

the first term being the Pearson estimator and the second term being a bias correction term. The corresponding adjusted Pearson estimator for v2 is

v1 =—^2 y y^ {0ij - Oi)2+a,+ct- iciWij i. m

i=l

Ji

j=l

l

}

Again, the first term is the Pearson estimator and the remaining terms are bias correction terms where Cy denotes the mean squared distance between the random effect Uij and its predictor, specifically dj = EiUij - Uij)2 = Wij (j/2 + CiWij) .

These dispersion parameter estimators can also be shown to be consistent as m —> oo. Unlike previous approaches in the literature, the asymptotic variance of our regression parameter estimator is not affected by variability in the dispersion parameter estimators. 4-4

Computational procedures

Due to the lack of closed-form solutions for those unknown parameters, the computational algorithms for random effects modeling are inevitably iterative. Initial values for the regression parameters are taken as the regression parameter estimates obtained from standard Poisson regression techniques assuming independent responses, Yijk,h- Initial random effects predictions Ui and Uij are given by the average of the responses within cluster i divided by the average of all responses, and the average of the responses within subject (i,j) divided by the average of all responses, respectively. The initial dispersion parameter estimates are calculated from the adjusted Pearson estimators, omitting the bias-correction terms. After initial values are given, the algorithm then iterates between updating the regression parameter estimates via the Newton scoring algorithm,

194

updating random effect predictors via the orthodox best linear unbiased predictor, and updating dispersion parameter estimates via the adjusted Pearson estimators until the algorithm converges. At each iteration, the left-hand sides in the above equations are updated by evaluating those unknown quantities on the right-hand sides at their current values.

5

An example

Fleming and Harrington 5 presented the chronic granulotomous disease (CGD) data of 128 patients from 14 medical clusters. These patients were randomized to either interferon gamma (rIFN-g, t r t = l ) or placebo (trt=0). Other variables such as age and sex were also recorded. The disorders were characterized by recurrent infections and the maximum number of infections observed was 7. Fourteen of the 63 patients in the treatment group had at least one infection and the total number of infections in this group was 20. On the other hand, 30 of the 65 patients in the placebo group had at least one infection and the total number of infections in the placebo group was 56. The question is if the treatment is effective in reducing the frequency of serious infections in patients. This problem is complicated by the possible intra-subject correlation and clustering effects of medical centres on these patients. To account for possible centre and subject effects, we consider four versions of the Cox models with both centre and subject level frailties specified by (1), (2) and (3). The treatment effect was taken as the only covariate. These four models are the standard Cox model without frailties (a2 = v2 = 0), centre frailty Cox model (i/2 = 0), subject frailty Cox model (a2 = 0) and the Cox model with both centre and subject frailties, respectively. The parameter estimates corresponding to these four models are displayed in the first, second, third and last row of table 1. This table shows that the dispersion parameter estimates for both centre and subject frailties in the fourth model are essentially zero. That is, there is little centre or subject effect. This conclusion has also been confirmed by the results based on the Cox models with a single level of frailties at either centre or subject level. This result is not surprising. In the literature, Therneau and Grambsch 14 found that there is little subject effect based on the Cox model with a single level of gamma frailties and Yau 16 found that there is little centre effect, but some subject effect based on the Cox model with both centre and subject level lognormal frailties.

195 Table 1. Parameter estimates for the chronic granulotomous disease data. Cox model Standard Center level Subject level Two levels

6

Treatment

fi

s.e.

-0.8759 -0.8759 -0.8759 -0.8771

0.2782 0.2782 0.2782 0.2785

Dispersion parameter Center Subject 0 0.0013

0 0.0009

Alternative Models

We analyzed the CGD data based on our model specified by (1), (2) and (3) where recurrent events times are stratified on the basis of event type, that is, the number of recurrences. More specifically, all the first event times form the first stratum, all the second event times form the second stratum, and so on. It implies that a patient's risk of the next infection might have been changed after he experienced his latest infection. If the occurrence of each infection has permanently compromised the immune system, the risk of the subsequent infections will be increased. However, some clinical scientists believe that the risk of recurrent CGD infection remains unchanged regardless of the number of previous infections (Therneau and Grambsch 1 4 ). The data can then be modeled by assuming that all the event times share the same baseline hazard as follows (Andersen and Gill 1): Kjk(t) = Xo(t)uij exp{{3xijk(t)},

(11)

where the centre and subject frailties are still specified by (1) and (2). We fit the CGD data with the above model specified by (1), (2) and (11) with treatment as the sole covariate. The centre and subject level dispersion parameter estimates are 0.097 and 0.545, respectively. This result indicates some subject effects and relatively small centre effects. The regression parameter estimates standard error for treatment effect are now —1.1183 and 0.30465 in comparison with -1.0829 and 0.26763 for the standard Cox model without frailty. The standard error has increased slightly. The difference between (3) and (11) is that the recurrent event times are stratified on the basis of the number of recurrences in the former model, but not in the latter one. So the latter model can be mathematically viewed as a special case of (3) by setting a = 1. However, there is some conceptual difference between these two models in terms of risk sets. Suppose a patient has experienced his (k + l)th recurrent event at time r , then every patient who is under observation at time r is at risk at that time in the latter model. But in the former model, among all subjects who are under observation at

196

time r, only those who have experienced their feth recurrent event, but have not yet experienced their (k + l)th recurrent event are at risk at time T. The extension of both models to include extra stratification based on certain external variables is straightforward. Suppose these models are extended to include extra stratification based on, say gender; therefore the former model is still in the form of (3), but with 2a strata based on both gender and event type being indexed by k = 1,... ,2a. On the other hand, the latter model is also in the form of (3), but with only 2 strata based on gender being indexed byfc = l,2. 7

Discussion

Yau 16 has recently proposed a best linear unbiased predictor approach to the Cox model with nested log-normal frailties. His approach was a generalization of the linear mixed model equations of Henderson 8 where the estimating equations are obtained by differentiating the joint partial likelihood with respect to the regression parameters and the frailties. It is an analytically tractable analog of best linear unbiased predictors; therefore the corresponding frailties predictors are still called best linear unbiased predictors (Yau 1 6 ). However, these frailty predictors actually correspond to the conditional mode of the frailty given the data, and are thus neither linear nor unbiased in general for non-normal distributions. In fact, these pseudo best linear unbiased predictors often lead to biased estimating equations and inconsistent parameter estimators. This approach remains controversial, especially for cases where the number of centres increases with the sample size (Ha et al. 6 ) . On the other hand, our approach to the nested frailty Cox model for recurrent events is based on a truly best linear unbiased predictor approach to the random effects Poisson models. Our approach gives unbiased estimating equations and leads to optimal and consistent parameter estimators. The extension of our approach to the Cox models with more levels of nesting effects is also straightforward. Acknowledgments This work was supported in part by the Natural Sciences and Engineering Research Council of Canada and in part by the Geomatics for Informed Decisions network of Canada. The authors gratefully acknowledge the computing support of Dr. Edward Hughes. The authors are also very grateful to Drs. Xuming He and Qiman Shao for their further clarification on their asymptotic results on parameters of increasing dimension.

197

References 1. P.K. Andersen and R.D. Gill, Ann. Statist. 10, 1100 (1982). 2. P.K. Andersen, J.P. Klein, K.M. Knudsen and P.R. Tabaneray, Biometrics 53, 1475 (1997). 3. R. Artes and B. J0rgensen, Scand. J. Statist. 27, 321 (2000). 4. D.R. Cox and D. Oakes, Analysis of Survival Data (Chapman and Hall, New York, 1984). 5. T.R. Fleming and D.P. Harrington, Counting Processes and Survival Analysis (Wiley, New York, 1991). 6. I.D. Ha, Y. Lee and J.K. Song, Biometrika 88, 233 (2001). 7. X. He and Q. Shao, J. Multi. Anal. 73, 120 (2000). 8. C.R. Henderson, Biometrics 3 1 , 423 (1975). 9. B. J0rgensen, S. Lundbye-Christensen, X.-K. Song and L. Sun, Statist. Med. 15, 823 (1996). 10. J.P. Klein, Biometrics 48, 795 (1992). 11. R. Ma, D. Krewski and R. Burnett, Random effects Cox model: a Poisson modelling approach (Tech. Rep. No 338. Laboratory for Research in Statistics and Probability, Carleton University, Canada, 2000). 12. G.G. Nielsen, R.D. Gill, P.K. Andersen and T.I.A. S0rensen, Scand. J. Statist. 19, 25 (1992). 13. N. Sastry, J. Am. Statist. Assoc. 92, 426 (1997). 14. T.M. Therneau, and P.M. Grambsch, Modeling Survival Data: Extending the Cox Model (Springer-Verlag, New York, 2000). 15. J. Whitehead, Appl. Statist. 29, 268 (1980). 16. K.K.W. Yau, Biometrics 57, 96 (2001).

MULTIPLE C O M P A R I S O N P R O C E D U R E S FOR LINEAR MODELS U N D E R O R D E R R E S T R I C T I O N S HOSSEIN MANSOURI AND ROBERT PAIGE Department

of Mathematics and Statistics, Texas Tech Lubbock, TX 79409-1042 E-mail: [email protected]

University

Multiple comparison procedures are some of the most frequently use statistical methods. Although, there exists an extensive amount of literature on multiple comparison procedures, few articles have addressed the problem of multiple comparisons under order restrictions. Currently available procedures are mainly developed for the one-way layouts and hence are not applicable to more complex designs such as unbalanced factorial designs or analysis of covariance models. To achieve these extensions, the focus of the present investigation is to develop simultaneous multiple comparison procedures for linear models under order restrictions of the parameters.

1

Introduction

Consider the linear model Y = X/3 + e,

(1)

where Y is an nx 1 observation vector, J i s a n n x p known matrix, (3 is apX 1 vector of unknown parameters, and e is an n x 1 vector of independent and identically distributed random variables having a normal distribution with mean 0 and variance a2 > 0. A common problem of interest is to make simultaneous inferences about components of a linear function C/3, where C is a r x p matrix with known elements. Focusing on simultaneous testing of the constrained parameters, suppose that we are interested in testing the hypothesis H0:C/3 = d

(2)

against the one-sided alternatives Hi — Ho, where Hi:CP>d

(3)

Without loss of generality it can be assumed that d = 0. Depending on dimensionality of C, we are dealing with a different class of tests. For instance, for testing subhypotheses against ordered alternatives in regression models or testing for main effects in factorial analysis of variance models, then r < p, and for simultaneous one-sided multiple comparison

198

199 procedures r > p. A third class of tests which is the main focus of this investigation, is the problem of simultaneous multiple comparisons under the assumption that the parameter space is constrained a priori. Although the first two problems are well developed, at least theoretically, not much is available on the third problem. Simultaneous tests for the order restricted (one-sided) hypotheses are well developed. Well-known tests are the likelihood ratio test, Wald's test, and Rao's efficient scores test. These are discussed in detail in Robertson et al. 10 . Shapiro 12 presented a unified approach by assuming the parameter space to be a closed convex cone and the observations follow a multivariate normal distribution with a known dispersion matrix. Mansouri 4 proposed a test based on the aligned ranking approach. One-sided simultaneous multiple comparison procedures for linear models, although not discussed explicitly can be derived from the results in Hochberg and Tamhane 3 and is outlined in Westfall et al. 13 . Simultaneous multiple comparison procedures under order restrictions of the parameters has not received much attention in the literature. The existing techniques are mainly developed for the one-way layouts under the assumption of simple order restriction of the means. Marcus and Peritz 8 proposed onesided Scheffe-type simultaneous confidence intervals for monotone contrasts in the means of normal distributions, Williams 14 proposed a Tukey-type range test for monotone contrasts for the balanced one-way layouts, Marcus 7 proposed a two-sided Scheffe-type simultaneous confidence intervals for monotone contrasts for the means of normal distributions. Hayter 2 argued that simultaneous inference for monotone contrasts does not extend to simultaneous pairwise comparisons and developed a one-sided studentized range technique under the assumption of simple order without the restriction on monotonicity of the contrasts. Recently, Mansouri and Shaw 6 proposed distribution-free simultaneous multiple comparison tests under the assumption of simple order for the balanced complete and incomplete randomized blocks. Mansouri 5 extended the latter results for repeated measures designs with incomplete observations. The main focus of this article is to develop a technique for simultaneous multiple comparisons in linear models. This technique provides a procedure that is applicable to balanced as well as unbalanced designs in analysis of variance and analysis of covariance models. In addition, we consider both the one-sided inferences without assuming monotonicity of the contrasts. In Section2, a brief summary of simultaneous one-sided inference in linear models is presented. In Section 3, simultaneous multiple comparison procedures are proposed and their distributions are discussed.

200

2

One-Sided Tests

In this section, a brief discussion of one-sided tests for subhypotheses, (q = r < p) and simultaneous one-sided multiple comparisons, (r > q,q < p) will be presented. 2.1

One-Sided Test for Subhypotheses

Let X = (Xi,X2) and P = {P'x, /3'2)', where Xx is n x q, X2 is n x p - q, j3\ and P2 are g x 1 and (p — q) x 1, respectively. Then the linear model in (1), can be written as Y = X1p1+X2p2

+e

(4)

Without loss of generality, the linear hypotheses in (2) and (3) can be written as H0 : f31 =0 and Hi : /31 > 0. Let

3 = (3'i, #>)' be the unrestricted maximum likelihood estimator (MLE) of /3. Let S=(Y-

X0)'(Y

- Xp) + ( 3 - P)'X'X(J3 - p)

Then the likelihood ratio statistic is equivalent to min S — min S PeH0 PeHi =

min(3 -/31)'Ar11(3i-/5i) PeH0 1 min(31-j31)'An1(31-/31)

-

/3€H,

= 3'1An13i - 0i - fiY*uX@i - Fx) ^Pl'A^Pl where P\ is the MLE of Pi under Hi and A n = (-X1-X1 — XXX2{X2X2)~ 2

Under Ho and known a , the test statistic X2=o-2pl'A^P*i has a ^-distribution given by

X2Xi)

.

201 9

where the weights Uj are nonnegative satisfying X^LQWJ = ^ arK * Xo = 0Computation of these weights depend on q, A n and the alternative hypothesis Hi. More details on o/j-'s are provided in Shapiro 12 sec. 5. For the one-way layouts, uij 's are called level probabilities and are discussed in detail in Barlow et al. 1. 2.2

One-Sided Multiple Comparisons

Again consider the linear model (4), and assume that we are interested in testing Hio and Hu where Hi0 : ctfi = 0 and Ha : d£x > 0, i = 1, • • • , r > « Let /3 = (JSJ, /3 2 )' be the unrestricted MLE of/3 and let S2 be an unbiased estimator of a2. Define the test statistics Ti = t@1/S(dikllCi)1/2,

i = l,---,r.

Then we reject H,o in favor of Hu — H^ if

where ca is upper a-th quantile of the distribution of max; T;. We note that Ti, • • • , T r have a joint r-dimensional t-distribution with dispersion matrix R = D _ 1 / 2 C ' A n C - D ~ 1 / 2 , where D = diag^AnCi, • • • , c^Ancv), see Westfall et al. (1999), chapter 5. 3

Multiple Comparisons Under Order Restrictions

Consider the linear model in (4) and assume further that the parameter space is subject to the constraint that (31 s K., where /C is a closed convex cone. The simple order restriction, considered in the cited literature in Section 1 is a member of /C. Suppose that we are interested in testing H^ against Hu where Hi0 : ci/31 = 0 and Hn : ci/31 > 0, i = 1 , . . . , r > q

202

Let P{ denote the orthogonal projection of /31 onto /C, i.e. /3j is the solution of min ( 3 i - ^ i )

An(3i-/3,).

Define the test statistics T* = cfijo

(c-AnCi)

, i = l,...,r.

Using the Union-Intersection principle of Roy (1953), we reject HIQ in favor of HiX - iifjo if T*>ca,

i = !,•••

,r

where ca is upper a-th quantile of the distribution of max; T*. By converting the simultaneous testing problem to simultaneous confidence region, we obtain simultaneous upper confidence limits with family confidence coefficient 1 — a given by c^^c^i-c^KAnC;)1/2,

i = l , . . . ,r

To find the critical values of the sampling distribution of max; T*, in most cases one has to resort to simulation techniques. This is a common practice in testing under order restrictions as well as situations where simultaneous multiple comparison techniques are used. This is the case because in both inferential approaches, except for the balanced analysis of variance designs, the sampling distribution of the resulting statistic is intractable, hence there is no choice but to resort to simulation. Similarly, if we are testing HM against the alternative Hii-.difa^O,

i = l,---

,r>q

we reject Hi0 if \T*\ > da/2, where da is the upper a-th quantile of the distribution of max;|T*|. Finally, a simultaneous confidence region with a family confidence coefficient 1 — a is given by cjfr € {<$[ ± ^ ^ ( ^ n c ) 1 / 2 } ,

i=

l,---,q.

To simulate the null distribution of the test statistics, the following algorithm may be used. 1. Generate E\, • • • ,en from a standard normal distribution.

203

2. Generate Y\, • • • ,Yn using the model Y = model under the Ho — f]t #;o-

which is the reduced

XR/3R+£

3. Calculate Tf, • • • , Tr* (\T;\, ••• , \T*\). 4. Calculate maxi<j< r maxi<;< r \T*\T* for the one-sided (maxi<;< r \T*\ for the two-sided) simultaneous confidence interval. 5. Repeat the process M times. 6. Find ca(da/2) the upper a-th (a/2-th)quantile m&xi
of

Example

Our dataset comes from Montgomery 9 ( pp. 170). Here, an engineer is studying the effect of cutting speed on the rate of metal removal in a machining operation. However, the rate of metal removal is also related to the hardness of the test specimen. Five observations are taken at each cutting speed. The amount of metal removed (y) and the hardness of the specimen (x) are shown in the following table: 1000

Cutting Speed (rpm) 1200 1400

X

V

X

68 90 98 77 88

120 140 150 125 136

112 94 65 74 85

X

y

165 140 120 125 133

118 82 73 92 80

y

175 132 124 141 130

W e assumed analysis of covariance model y{j = (j, + Ti + 7 (xij -x..)

+ Eij for i = 1, 2,3 and j = 1 , . . . ,5

where e^ ~ N (0,CT2). Since the rate of metal removal can be expected to be monotonically increasing in cutting speed it is conceivable that Ti <

T2

< T3.

We then considered the following sets of hypotheses: #10 : r 2 - ri = 0 #20 : r 3 - n = 0 #30 : T 3 - r 2 = 0

#11 : r 2 - n > 0 #21 : r 3 - ri > 0 #31 : r 3 - T 2 > 0

(5)

204

1.6-

0.5 -

0.0

J

/

f^

1

1

1

1

0.0

0.6

1.0

1.5

2.0

'

Figure 1: Density Plot of max; T* Next, identifiability constraint ^ ; T; = 0 was imposed. Under this, constraint our omnibus null hypothesis was H0:TI=T2=T3=

0.

For the reduced model, under the omnibus null, we obtained the following parameter estimates: 7 = 0.9299 fi = y = 86.40 a = 2.744 We simulated from the estimated reduced model. Statistics Z\*, T% and T£, associated with following hypotheses sets (Hw, Hn), (H20, #21) and (H30, #31) respectively, were computed and the quantity max;T* was obtained. This process was repeated 9,999 times to yield a total of 10,000 simulated max; T* values. Next, ca, an estimate for ca, the upper a-th quantile of the max;T*, was obtained by first ordering the max;T/ values from smallest to largest and taking our estimate to be the realization of the 10,000 (1 — a)-th order statistic. Note that, for our example a was set at 0.05 which guaranteed an integral 10,000 (1 — a) value. Our simulated 0.05-th quantile of the max; T* turned out to be 1.1652. Figure 1 shows the S-Plus density plot of the simulated max; T* values.

205

For our example, the observed values of the T*s were as follows:

i

2?

1 2 3

0.4157 0.4237 0.2638

resulting in failure to reject Hio for i = 1,2,3. 5

Conclusion

In this article, we have proposed a method for making simultaneous one-sided and two-sided multiple comparisons for linear models, when the parameters of interest belong to a closed convex cone. The exact distributions for the resulting statistics are derived. However, because of complexity of the distributions for all except the balanced analysis of variance designs, one must obtain the critical values by simulation. An algorithm for generating the approximate values of these quantiles is proposed. Lastly, we applied this procedure to a data set. References 1. R.E. Barlow, D.J. Bartholomew, J.M. Bremner and H. Brunk, Statistical inference under order restrictions (John Wiley, New York, 1972). 2. A.J. Hayter, Journal of the American Statistical Association 85, 778 (1990). 3. Y. Hochberg and A.C. Tamhane, Multiple comparison procedures (John Wiley, New York, 1987). 4. H. Mansouri, Journal of Statistical Planning and Inference 24, 107 (1990). 5. H. Mansouri, Distribution-free multiple comparisons for incomplete data (Submitted for publication, 2000). 6. H. Mansouri and C. Shaw, Nonparametric multiple comparison procedures for ordered parameters in balanced incomplete blocks (Submitted for publication, 1999). 7. R. Marcus, Communications in Statistics-Theory and Methods 1 1 , 615 (1982). 8. R. Marcus and E. Peritz, Journal of the Royal Statistical Society, Ser. B 38, 157 (1976). 9. D. Montgomery, Design and Analysis of Experiments, 4th edition (John Wiley, New York, 1997).

206

10. T. Robertson, F.T. Wright and R.L. Dykstra, Order restricted statistical inference (John Wiley, New York, 1988). 11. S.N. Roy, Annals Of Mathematical Statistics 24, 220 (1953). 12. A. Shapiro, International Statistical Review 56, 49 (1988). 13. P.H. Westfall, R.D. Tobias, D. Rom, R.D. Wolfinger and Y. Hochberg, Multiple comparisons and multiple tests using the SAS system (SAS Institute, Cary, NC, USA 1999). 14. D.A. Williams, Biometrika 64, 9 (1977).

TESTING GOODNESS-OF-FIT OF THE G A M M A MODELS

C A R O L E. M A R C H E T T I Department

of Mathematics and Statistics, Rochester Institute Rochester, NY 14623, USA E-mail: [email protected]

of

Technology,

G O V I N D S. M U D H O L K A R A N D G R E G O R Y E . W I L D I N G Department

of Biostatistics, University of Rochester, Rochester, NY 14642, E-mail: [email protected]; [email protected]

USA

Gamma models are extensively employed in statistical analyses, especially for reliability and lifetime data. Assessing the appropriateness of these models is of obvious importance. While characterization theorems in probability and statistics are widely appreciated for their role in clarifying the structures of families of probability distributions, they are not as well recognized for being natural, logical and effective starting points for constructing tests for goodness-of-fit problems. It is well known that two independent identically distributed random variables ( X i , X2) have a gamma distribution if, and only if, X i + X% and X1/X2 are independent. This characterization is used by Locke 1 8 to propose a method for testing the composite gamma hypothesis. It is based upon the random creation of n pairs from a sample of size 2n, and using tests of bivariate independence, such as those based on the rank correlation coefficient or Kendall's tau, for testing independence of the sums and the ratios. However, the tests of bivariate independence he considered are consistent against only some dependence alternatives. Instead, we propose employing the classical rank test due to Hoeffding 9 or its asymptotic equivalent due to Blum, Kiefer and Rosenblatt 2 , which are known to be consistent against all dependence alternatives. The resulting goodness-of-fit tests are consistent against all non-gamma alternatives and, in moderate size samples, offer substantial power advantages. Additionally, an alternative approach, which does not suffer from the caprice of chance implicit in the random pairing, is also considered.

1

Introduction and Summary

Characterization theorems in probability and statistics are generally well appreciated for their aesthetic appeal, mathematical completeness and the light they shed on the structures of the probability distributions. Although logically self evident, but not well recognized, is the fact that they can be natural and effective bases for constructing goodness-of-fit (GOF) tests needed for assessing the validity of models based on parametric families, such as normal, exponential, and inverse Gaussian (IG), commonly used in statistical practice. The earliest explicit use of a characterization theorem for constructing a goodness-of-fit test is by Vasicek 37 , who used Shannon's maximum entropy characterization to construct a test for the composite hypothesis of normality.

207

208

Now there exists a substantial body of literature on the goodness-of-fit tests based on characterization theorems. Maximum entropy characterizations have been used by Gokhale 7 and Mudholkar and Lin 27 to construct goodness-offit tests for exponentiality, and more recently by Mudholkar and Tian 30 to construct a GOF test for the inverse Gaussian model. Characterization results based on the statistical independence of sample statistics have been used by Lin and Mudholkar 15 and Mudholkar, Marchetti and Lin 2 8 to construct GOF tests for normality, and by Mudholkar, Natarajan and Chaubey 29 to test inverse Gaussian goodness-of-fit. For an overview of characterizations and goodness-of-fit, see Marchetti and Mudholkar 20 . The gamma family of unimodal, right-skewed distributions with nonnegative support, has been used to model a wide variety of applications in diverse fields such as geology (McCullagh and Lang 2 5 ) , ecology (Matis, Rubink, and Makela 2 3 ) , inventory control and queuing (Yeh 4 0 ) , economics (McDonald and Jensen 2 6 ) , meteorology (Bougeault 3 ) , reliability (Reiser and Rocke 3 5 ) , biomedical studies (Tan 36 ) and genetics (Yang 3 9 ) . Testing the appropriateness of the gamma models for use in these applications is vital. Empirical distribution function (EDF) tests for the composite gamma hypothesis have been developed by Pettitt and Stephens 34 and Lockhart and Stephens 16 . For an account of these, see D'Agostino and Stephens 4 . It is well known that the distribution of two IID random variables (X\,X2) is gamma if and only if X\ + X2 and X\jX2 are independent (see Lukacs, 1 9 ). Locke 17,Locke76 proposed testing the composite gamma hypothesis by creating n pairs from a sample of size 2n and testing the independence using procedures such as the one based on Kendall's tau. Since the rank tests of bivariate independence he considered are consistent against only some dependence alternatives, the resulting gamma GOF tests lack consistency against all non-gamma alternatives. The existing goodness-of-fit tests of the composite gamma hypotheses are reviewed in Section 2. A modification of one of these tests, together with a comparative study of its power function is presented in Section 3. The modified test, which is consistent against all non-gamma alternatives, offers a power advantage with moderate sample sizes. In Section 4, an alternative approach is considered. Section 5 is given to conclusions. 2

The Gamma Goodness-of-Fit Tests

The literature on testing goodness-of-fit for the gamma models is rather scant, and most of the proposed solutions lack simplicity for use in practice. The tests available, which may be grouped as the EDF tests and the tests based on characterization results, are now reviewed.

209 2.1

The Empirical Distribution Function Tests

The empirical distribution function for a sample Xi,X%, ...,X„ is defined by Fn(x) = # ( X j < x)/n. To test that the sample is from a population with a distribution function F(x) = F0(x), where F0(x) is fully specified, one may measure the discrepancy between the Fn(x) and F0(x). Kolmogorov 13 introduced the first EDF statistic, D, by defining D+ = sup{Fn{x) - F0{x)},

(1)

X

£>_ = sup{F0(a;) - Fn(x)},

and

(2)

X

D = sup \Fn{x) - F0(x)\ = max(D + , DJ).

(3)

X

The Cramer-von Mises family is a class of statistics of the form oo

/

{Fn(x)-F0{x)}21>{x)dF[x).

(4)

-oo

When ip(x) = 1, the Cramer-von Mises W2 statistic is obtained; when i[>(x) = [F(x)(l — F(x))}~1, one obtains the Anderson-Darling * statistic A2. The EDF tests have been adapted for the composite goodness-of-fit hypothesis F(x) = F0(x, 6), where F0 is known but 6 is not, by plugging in estimated 0 for 6. Pettitt and Stephens 34 considered the problem of testing the composite gamma hypothesis in which the shape parameter a is known and scale parameter /? is unknown. They provide asymptotic percentiles for three EDF statistics for the problem. Lockhart and Stephens 16 examined the situation of a gamma distribution with unknown shape parameter and scale parameter either known or unknown. They suggest using a maximum likelihood estimator of a and give asymptotic percentiles for the resulting EDF statistics. D'Agostino and Stephens 4 provide an excellent overview of goodness-of-fit for the gamma and other distributions. 2.2

Characterizations and Gamma GOF Tests

A nice summary of characterizations of the gamma family, from the earliest results due to Nabeya 3 3 , Goodman 8 , Mauldon 2 4 and Laha 14 , appears in Johnson, Kotz and Balakrishnan 12 . The simplest of these, the following result due to Lukacs 19 , is the basis of Locke's 18 test. Proposition 2.1: Two independent random variables (X\,X2) have a common gamma distribution iff X\ + X-i and X\jXi are independent.

210

Extensions, applications and reviews of the above characterization may be found in Findeisen 6 , Marsaglia 21,22 and Wang 38 . Since this characterization is asymmetric in Xt and X2, Locke 17 considers the dependence in terms of U = Xi + X2 and V = ma.x(X1/X2, X2/X1). Before discussing the specifics of Locke's test we note a general maximum entropy result, from which characterizations of several members of the exponential family have been fruitfully used to develop consistent goodnessof-fit tests. The examples include the maximum entropy tests of the composite hypotheses of normality (Vasicek 3 7 ) , uniformity (Dudewicz and van der Meulen 5 ) , exponentiality (Gokhale 7 , Mudholkar and Lin 27 ) and the inverse Gaussian distribution (Mudholkar and Tian 30>31). Gokhale 7 presents a general discussion of construction and suggests a GOF test for the gamma model, but offers no details. The following is a relatively recent characterization due to Hwang and Hu n which will be considered in Section 4. Proposition 2.2: The mean x and coefficient of variation s/x of a random sample are independent iff the population is gamma. 2.3

Locke's Tests

Specifically, Locke 18 proposed testing the composite gamma hypothesis by randomly pairing observations from a sample of size 2n, to obtain n pairs (Yi,Zi), where Yi = X2i + X2i-i,

Zi = m&x{X2i/X2i-i,X2i-i/X2i},

(5)

i = 1,2, ...,n. Then testing the composite gamma hypothesis is equivalent to testing the independence of Y and Z. Locke employs the quadrant tests and tests based on the rank correlation coefficient and Kendall's tau for this purpose. An obvious drawback to this method is that with random pairing, the same data set can produce different results. However, more importantly the rank tests that Locke considers are consistent against only some dependence alternatives. In the next section, we propose a modification of Locke's procedure that addresses the consistency issue. 3

A Consistent Modification of Locke's Procedure

Locke's 18 approach of randomly pairing the observations in the sample can be combined with use of the consistent tests of bivariate independence due to Hoeffding 9 and Blum, Kiefer and Rosenblatt 2 to construct gamma GOF

211

tests which would be consistent against all non-gamma alternatives. These modifications would be considered improvements over Locke's test provided they show power advantages with moderate size samples. In this section, we first describe the consistent tests of bivariate independence, the modification of Locke's procedure, and then present the results of a study of the power functions. 3.1

Two Consistent Tests of Bivariate Independence

Hoeffding's

9

notes that D(x, y) = FX,Y ( * , » ) - FX (x) FY (y) = 0,

(6)

if and only if X and Y are independent. Nonparameteric estimation of the quantity J D2(x, y)dF(x, y) results in the statistic,

_Q-2(n-2)R+(n-2)(n-3)S n(n-l)(n-2)(n-3)(n-4)

'

^'

in which n

Q = J2(Ri - !)(*i - 2)(Si ~ !)(£ - 2),

(8)

1=1 71

22 = 5^(i2i-2)(5 i -2)c i ,

s

i=i n

= I > < - 1)Ci-

(9)

(10)

t=i

where Ri and Si are the respective ranks of Xi among the A"'s and Yi among the y ' s , and Ci is the number of bivariate observations (Xj,Yj) for which Xj < Xi and Yj < Yi. Hoeffding 9 provides the exact null distribution of nDn for sample sizes n = 5,6,7 which is extended to n = 8,9 by Hollander and Wolfe 10 . However, Hoeffding 9 was unable to obtain the asymptotic distribution of nDn. Blum, Kiefer and Rosenblatt 2 consider a computationally simpler statistic, n

Bn = n- 5 J2 Wi(i)N4(i) - N2(i)N3(i)f,

(11)

t=i

where N\(i), N2(i), Ns(i), and JV4(i) are the number of points lying in the four quadrants determined by the vertical and horizontal lines through the

212

bivariate point (Xi, Yj). Independence is rejected for large values of the test statistic. Blum, Kiefer and Rosenblatt 2 show that nBn is asymptotically equivalent to Hoeffding's 9 statistic and provide the asymptotic percentiles. Mudholkar and Wilding 32 obtain the empirical distribution of the two statistics for small and moderate size samples. A selection of the 10%, 5% and 1% points of the nDn and nBn statistics appears in Table 1. A Gaussian approximation for the nBn statistic, which can be used to obtain both the associated percentiles and probabilities, is currently under development by Mudholkar and Wilding. 3.2

The Modified Test

The tests of bivariate independence due to Hoeffding 9 and due to Blum, Kiefer and Rosenblatt 2 may be used for testing the independence of Y and Z. Since the two rank tests are consistent against all dependence alternatives, it is obvious that the modified versions of Locke's procedure using these tests will produce goodness-of-fit tests which are consistent against all non-gamma alternatives. Let Xi,X2,..., Xin be a random sample from a population with d.f. F and consider the problem of testing the composite hypothesis that F(-) is the d.f. of a gamma random variable. If the sample size is odd, as in Locke 18 , delete an observation at random. We propose testing the composite hypothesis by randomly pairing observations from the sample, to obtain n pairs (Yi, Zi), as given in (5). Testing the composite gamma hypothesis is equivalent to testing the independence of Y and Z. The rank tests due to Hoeffding 9 and to Blum, Kiefer and Rosenblatt 2 are employed for this purpose. 3.3

A Monte Carlo Experiment

A Monte Carlo experiment was conducted using 100,000 simulated samples of size n = 20,30,50,100 from a variety of non-negative populations. Each sample was subjected to the above tests of the composite gamma hypothesis which are based on Locke's premise and use tests of bivariate independence due to Kendall, Spearman, Hoeffding and Blum et al. A selection of the results of the simulation experiment are shown in Table 2. In the population column, GTL stands for Generalized Tukey-Lambda family. Where parameters are unspecified, the standard distribution is used. The Spearman, Hoeffding and Blum et al. tests have excellent Type I Error control, but the Kendall test is not so accurate in this respect for small or even moderate sample sizes. For most of the populations, but not all, the tests

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4

A n Alternative Approach

To remove the caprice inherent in the random pairing of observations, another approach to testing goodness-of-fit for the gamma distribution is in progress. It is based upon the following recent characterization of the gamma distribution due to Hwang and Hu u , given in Proposition 2.2. For developing the test we use the technique introduced by Lin and Mudholkar 15 in their test for normality. Given a random sample x\, X2, • ••, xn, we create n pairs (£_i, c_;) by removing one observation at a time from the sample. The product moment correlation between the pairs is used to measure dependence between x and c. We therefore consider the statistic r =

EiU(5-i-g)(c-i-c) > / £ " = i ( 2 - i - S) 2 E L i ( c - i - c) 2

( 12 x

where c = n _ 1 £ " = 1 c_j. The asymptotic null distribution of r, however, depends on the shape parameter of the gamma distribution a. y/nr-+N(o,3+™).

(13)

One can compute the maximum likelihood estimate & and substitute to obtain a measure of the standard error, but such estimation requires solving a nonlinear equation involving the di-gamma function. There is also the question of use with small samples. Investigation of these issues is in progress. 5

Conclusion

In this paper we have examined the use of Hoeffding's 9 test of bivariate independence and its asymptotic equivalent due to Blum, Kiefer and Rosenblatt 2 , to obtain consistent modifications of Locke's test of the composite gamma hypothesis. It is seen that the modified tests have substantial power advantages over the original. References 1. T.W. Anderson and D.A. Darling, Journal of the American Statistical Association 49, 765 (1954). 2. J. R. Blum, J. Kiefer and M. Rosenblatt, The Annals of Mathematical Statistics 32, 485 (1961). 3. P. Bougeault, Journal of the Atmospheric Sciences 39, 2691 (1982).

216

4. R.B. D'Agostino and M.A. Stephens, Goodness-of-fit Techniques ( Marcel Dekker, New York, 1986). 5. E. Dudewicz and E.C. van der Meulen, Journal of the American Statistical Association 76, 967 (1981). 6. P. Findeisen, Annals of Statistics 6, 1165 (1978). 7. D.V. Gokhale, Computational Statistics and Data Analysis 96, 157 (1983). 8. L.A. Goodman, Annals of the Institute of Statistical Mathematics 3, 123 (1952). 9. W. Hoeffding, The Annals of Mathematical Statistics 19, 546 (1948). 10. M. Hollander and D.A. Wolfe, Nonparametric Statistical Methods (John Wiley, New York, 1999). 11. T.Y. Hwang and C.Y. Hu, Annals of the Institute of Statistical Mathematics 5 1 , 749 (1999). 12. N.L. Johnson, S. Kotz and N. Balakrishnan, Continuous Univariate Distributions (John Wiley, New York, 1994). 13. A. Kolmogorov, Gior. 1st. Ital. Attuari 4, 83 (1933). 14. R.G. Laha, Annals of Mathematical Statistics 25, 784 (1954). 15. C.T. Lin and G.S. Mudholkar, Biometrika 67, 455 (1980). 16. R.A. Lockhart, and M.A. Stephens, Goodness-of-fit tests for the gamma distribution. (Technical Report, Department of Mathematics and Statistics, Simon Fraser University, 1985). 17. C. Locke, Communications in Statistics - Theory and Methods 4, 357 (1975). 18. C. Locke, Communications in Statistics - Theory and Methods 5, 351 (1976). 19. E. Lukacs, The Annals of Mathematical Statistics 26, 319 (1955). 20. C.E. Marchetti and G.S. Mudholkar, Goodness-of-Fit Tests and Model Validity (Birkhauser, Boston, 2001). 21. G. Marsaglia, Proceedings of the Symposium on Statistics and Related Topics (Carleton Univ., Ottawa, Ont., 1974). 22. G. Marsaglia, In Contributions to Probability and Statistics. Essays in Honor of Ingram Olkin (Springer-Verlag, New York, 1989). 23. J. H. Matis, W. L. Rubink and M. Makela, Environmental Entomology 21, 436 (1992). 24. J.G. Mauldon, Quarterly Journal of Mathematics 27, 155 (1956). 25. P. McCullagh and P. Lang, Journal of the Royal Statistical Society B 46, 344 (1984). 26. J.B. McDonald and B.C. Jensen, Journal of the American Statistical Association 74, 856 (1979).

217

27. G.S. Mudholkar and C.T. Lin, Colloquia Mathematica Societatis, Janos Bolyai 45, 395 (1984). 28. G.S. Mudholkar, C.E. Marchetti and C.T. Lin, Jounal of Statistical Planning and Inference (To appear, 2002). 29. G.S. Mudholkar, R. Natarajan and Y. P. Chaubey, Sankhya Ser. B 63, 362 (2002). 30. G.S. Mudholkar and L. Tian, in Jounal of Statistical Planning and Inference (To appear, 2002). 31. G.S. Mudholkar and L. Tian, Communications in Statistics - Theory and Methods (To appear, 2002). 32. G.S. Mudholkar and G.E. Wilding, On the conventional wisdom regarding two consistent tests of bivariate independence (Technical Report, Department of Biostatistics, University of Rochester, 2001). 33. S. Nabeya, Annals of the Institute of Statistical Mathematics 2, 13 (1950). 34. A.N. Pettitt and M.A.Stephens, EDF Statistics for Testing the Gamma Distribution (Technical Report, Department of Statistics, Stanford University, 1983). 35. B. Reiser, and D.M. Rocke, IAPQR Transactions, Journal of the Indian Association for Productivity, Quality & Reliability 18, 1 (1993). 36. W. Y. Tan, Biometrical Journal 37, 319 (1995). 37. O. Vasicek, Journal of the Royal Statistical Society B 38, 54 (1976). 38. Y.H. Wang, Analytical Methods in Probability Theory, Lecture Notes in Math. 861, 166 (Springer, New York, 1981). 39. Z. Yang, Journal of Molecular Evolution 39, 105 (1994). 40. Q. Yeh, Production and Inventory Management 38, 51 (1997).

O N FRAILTY MODELS A N D C O P U L A S DAVID OAKES Department of Bio statistics, University of Rochester Medical Center 601 Elmwood Avenue Box 630, Rochester NY 14642, USA E-mail: [email protected] We review the connection between archimedean copula models and frailty models for bivariate failure-time data, emphasizing applications in biostatistics, reliability and extreme value theory. A new method for semiparametric estimation of the dependence parameter in an archimedean copula model is proposed.

1

Bivariate Frailty Models

Many models have been proposed for multivariate data failure-time data (Ti,T 2 ) arising in biostatistics, reliability and other applications. The classical model of Marshall and Olkin 16 has a discontinuity along the diagonal Ti = Ti which may be appropriate if the Tj represent times to failure of a machine component and failure of one component results in increased stress on the other component. However this model would not be appropriate for the data reported in Table 1 of Oakes 21 from Lawless 12 (p. 477) in which Ti is the time to appearance of a fracture in a component and T2 is the subsequent time to failure. A plot of the data, shown in Oakes 21 (Figure 2) suggests that Ti and T2 are not independent so that a model is needed which allows the bivariate survivor function pr(Ti > t\,Ti > £2) = S{t\,t-i) to be continuous over the whole positive quadrant while still allowing dependence between T\ and T 2 . Oakes 22 analyzed an example of Gumbel and MustafA 8 from extreme value theory. Here T\ is the height in inches of the flood of the Fox River at Berlin, WI (upstream) and T2 the height at Wrightstown WI (downstream). Of the many possible models we focus on one class, bivariate frailty models, which allow for very flexible modeling of the marginal distributions of T\ and T2 and, separately, of the dependence between them. We assume that there is an unobserved random variable W, called here a frailty, common to Ti and T 2 , and such that Ti and T2 are conditionally independent given W. In the reliability example above, W might represent the susceptibility or inherent weakness of the component. In studies of human populations W might represent the influence on mortality of unmeasured genetic or environmental factors common to two related individuals. Each of Ti and T2 follows a proportional hazards model in W, so that Pv(Tj>tj\W

= w) =

218

Bj(tJr,

219

where the Bj(tj) are baseline survivor functions, corresponding to a component with unit frailty W = 1. The law of iterated expectations gives S(tut2)

= E{pr(T! > tuT2 > h\W)}

=

E{B1{tl)wB2{t2)w)

= p[-log{fli(ii)}-log{B 2 (i 2 )}]. where p{-) = E { e x p ( - • W)} is the Laplace transform of the distribution of W. Often the Bj(tj) will not be observable. However if we write q(-) for the inverse function to p(-) we obtain the expression S(h,t2)

=p[q{S1(t1)}

+q{S2(t2)}}

for S(ti,t2) in terms of its marginal survivor functions Si(ti) = S(ti,0) and 52(42) = S(0,t2). Genest and MacKay 5 described these as "archimedean copula models". When W follows a gamma distribution with unit scale parameter and index K, we have p{s) = (1 + s)~K and

S{tut2)

~ [l-log{Bi(*i)}-log{B2(t2)}J .5i(«i)(- 1 /*) +S2(t2)(-i/*)

-1_ '

When Bj(tj) = exp(-pjtj) we obtain the bivariate Pareto distribution S(h,t2) = (1 4- pih + p2t2)~K. The form of S{ti,t2) with exponential marginals is mentioned in Johnson and Kotz u (p. 288), but the general form appears to have been first proposed by Clayton 2 , who also gave an alternative derivation via local odds ratios that is described below. The model was later given independently by Lindley and Singpurwalla 14 who viewed the W as the equivalent of a "random environment". By allowing W to have the so-called "positive stable distribution" (Feller, 3 Chapter XII) with Laplace transform p(s) — exp(—s a ), for some 0 < a < 1, we obtain a family considered by Gumbel, 7 with S(tut2) Hougaard 23

9

= exp{-([-log{5 1 (t 1 )}]( 1 / a ) +

{-\og{S2(t2)}]^r}.

gave the derivation via frailties. See also Oakes and Manatunga

220

2

Local Odds ratios

The question arises as to whether the distribution of W is determined by 5(^1,^2)- Oakes 22 answered this question in the affirmative, by considering the local odds ratio

Here the superscripts denote order of differentiation in h and t 2 so that the terms in the numerator are the joint density and survivor functions of (Ti, T2) and those in the denominator are the two mixed survivor-density functions. The function 9*(ti,t2) may be thought of as the odds ratio of the 2 x 2 table formed by dichotomizing each side of the region above and to the right of the point (ti,t2) at ti + Ai and t2 + A 2 respectively, where the Aj are small. It also equals the ratio of the conditional hazard functions at i 2 of T2 given Ti = *i and of T2 given Ti > h. It is easily seen that

p'{s)z where s = q{Si(ti)} + q{S2(t2)} = q{S(h,t2)}, and so depends on (ti,t2) only through the value v say of S(t\,t2). That is 6*{t-\_,t2) — 0{S{t\,ti)} for some function 9(v). Moreover, since s = q(v), the inverse function theorem gives

q'(v) which can be integrated twice, showing that 9(v) determines q(v) up to an arbitrary multiplicative factor. Hence p(s) is determined up to a scale factor by 9(v). For Clayton's model, with a gamma frailty distribution, we find that 9(v) = 1/K + 1, i.e. free of (^1,^2), while for Hougaard's model, 9{v) = 1 +

1 a ~ a log(w)'

which declines from infinity to unity as v declines from v — 1 to v = 0. Hougaard's model seems more realistic in most applications.

221 3

The Kendall Distribution

An alternative approach to characterizing archimedean copula models, which include frailty models, was proposed by Genest and Rivest 6 . They considered the (univariate) distribution function K(v) and density k(v) of the random variable V = S(T\,T2), analogous to the usual univariate probability integral transform. Here the distribution of V is not uniform, but has V

'

«'(«)'

V

'

q'(v)2

The similarity of the functional forms of 9(s) in terms of p(s) and k(v) in terms of q(v) is purely coincidental. It is clear however that either K{v) or k(v) determines q(v) up to a constant factor. For Clayton's model we find that K(V)

= (K +

1)V-KV1+1'K,

and for Hougaard's model K(v) = v(l — alogv). For Hougaard's model it turns out that Z — — log(F) has the simple mixed gamma density f(z) = (1 - a + az)exp(-z), a result originally given by Lee 13 . An immediate corollary of Lee's result is that if Z has the mixed gamma density f(z) and U is uniform (0,1), with Z and U independent, then T = UaZ has a unit exponential distribution. This amusing result can be shown directly from the facts that E{Tn) = E(Una)E{Zn)

= —^—-{(1 - a)n! + a(n + 1)!} = n\, na + 1 and that the exponential distribution is determined by its moments (Feller, 3 p. 234). To see the reason for the name Kendall distribution, note that the population analog r of Kendall's coefficient of concordance f can be written T = 4pr(Tj' > TUT^ > T2) - 1 where the pairs (TUT2) and (T{,T£) are drawn independently from S(ti,t2). We have r = 4£{pr(T 1 ' > TUT!, > T2)\TUT2)

- 1 = 4£S(T 1 ,T 2 ) - 1 = 4E(V) - 1.

For bivariate frailty models r = 4 J sp"(s)p(s) r = 1/(1 + 2K), while Hougaard's model has r = Given a random sample {(T{1>,Tj,1'), i = ate distribution, a simple analog estimator K(v)

- 1. Clayton's model has 1 - a. l , . . . , n } from the bivariof K(v) can be calculated

222

as the empirical distribution function K(v) = n _ 1 # { i : Vi < v], where Vi = (n + l)-x#{j : T^ > T^,TP > T 2 W }. Plots of K(v) against v or possibly K(v)/v - 1 against v may be useful diagnostics. Barbe et al. x prove weak convergence of a suitably normalized version of Kendall's process {K(v), 0 < J) < 1} to a Gaussian process and give an expression for the covariance function. Note that the limiting covariance exhibits long-range dependence. A smoother plot may be obtained from risk sets formed from the componentwise minima of two pairs of observations, i.e. from the n(n - l ) / 2 pairs ( T W ) i r W ) ) w h e r e T™ = m i n ( T ^ T ^ ) and T^'j) = min(T 2 W ,T 2 0 ) ). Let Rij denote the size of the corresponding bivariate risk set, i.e. Rij =#{k:

T[k) > T^j\T^]

> Tfj)}.

and

Aij=sign{(T^-T^)(T^-T^)}, the indicator of concordance or discordance for the pair (i,j). that nr./A

_ i |T(«.J) _

prCA^-lIT,

f

-tl,T2

T(i,3)

_ 4.

D

_

r

\ _

It is easily seen

0*(*1,*2)

-h,B*j-T)-r_l

+

e.{hM).

This approach relates closely to the original proposal of Clayton (1978) for estimation of 6 in his model: here the conditional probability that a pair is concordant given the size of the bivariate risk set is simply p

p r ( A y = l\Rtj =r) = r _

1 + e-

For general archimedean copula models pr(Ay = l\Rij = r) Oakes 4

22

6(r/n) r - 1 + 6(r/n)'

gave the exact probability for the positive stable frailty model.

Estimation of a Dependence Parameter

Suppose that a copula model is specified up to a single parameter a indexing the dependence and the marginal distributions are either parameterized fully or are kept arbitrary. Then the following strategies are possible for estimating Q.

223

(a) (Semi)parametric maximum likelihood or an equivalent technique. Results of Murphy 17 ' 18 and Parner 24 for special frailty models and of Murphy and van der Vaart 19 for general semiparametric profile likelihood functions suggest that this approach can be expected to be consistent, asymptotically normal, and asymptotically fully efficient. However there are some questions about its performance in small samples. The EM algorithm can sometimes be used to assist computation. (b) Marginal profile ("pseudo-")likelihood estimation. Fit each marginal separately, either by parametric or nonparametric maximum likelihood, and then assume that each marginal is fixed at its estimated value and maximize the likelihood in a. See, for example, Shih and Louis, 25 Genest, Ghoudi and Rivest 4 and Hougaard 1 0 ) . This approach appears to work well in practice, has a relatively simple asymptotic theory, but is not fully efficient. (c) Use an analog estimator based on Kendall's r or another similar summary measure of correlation (Oakes, 20 Manatunga and Oakes 15 and Genest and Rivest 6 ) . (d) Derive estimating equations based on bivariate risk sets, i.e. consider pr(Ay = l\Rij) - see the discussion at the end of the previous section. (e) Fit the Kendall density k(v) to the distribution of the empirical estimates Vi defined in Section 3. Asymptotic theory needs to be modified to account for the correlations among the Vi. Properties of this estimator are being studied by A. Wang and D.O. It seems to work quite well. Details will be given elsewhere. We used this approach to compute estimates K, and d and the corresponding values f of Kendall's tau from fitting the gamma and positive stable frailty models to the data for the two examples mentioned in Section 1. For the cable insulation data (with f = 0.46) the estimates from the gamma model and positive stable model are respectively Gamma Model: h = 0.46 f = 0.52, Stable Model: a = 0.45 f = 0.55 . For the Fox river data (with f = 0.52) we follow Gumbel and Mustafi 8 and Oakes 22 in applying our models to the inverted ranks, which gives, Gamma Model: k = 0.38 f = 0.57, Stable Model: a - 0.43 f = 0.57 .

Acknowledgments This work was supported by grant R01 52572 from the (US) National Cancer Institute.

224

References 1. P. Barbe, C. Genest, K. Ghoudi and B. Remillard, J. Mult. Anal. 58, 197 (1996). 2. D. G. Clayton, Biometrika 65, 141 (1978). 3. W. Feller, An Introduction to Probability Theory and its Applications, Vol. II, 2nd Ed. (John Wiley, New York, 1971). 4. C. Genest, K. Ghoudi and L.-P. Rivest, Biometrika 82, 543 (1995). 5. C. Genest and R. J. MacKay, Can. J. Statist. 14, 145 (1986). 6. C. Genest and L.-P. Rivest, J. Am. Statist. Assoc. 88, 1034 (1993). 7. E. J. Gumbel, J. Am. Statist. Assoc. 55, 698 (1960). 8. E. J. Gumbel and C.K. Mustafi, J. Am. Statist. Assoc. 62, 569 (1967). 9. P. Hougaard, Biometrika 73, 671 (1986). 10. P. Hougaard,Analysis of Multivariate Survival Data (Springer-Verlag, New York, 2000). 11. N. L. Johnson and S. Kotz, Continuous Univariate Distributions (John Wiley, New York, 1972). 12. J. Lawless, Statistical Models and Methods for Lifetime Data (John Wiley, New York, 1982). 13. L. Lee, J. Mult. Anal. 9, 266 (1979). 14. D.V. Lindley and N.D. Singpurwalla, J. Appl. Prob. 23, 418 (1986). 15. A.K. Manatunga and D. Oakes, J. Mult. Anal. 56, 60 (1996). 16. A. W. Marshall and I. Olkin, J. Am. Statist. Assoc. 62, 30 (1967). 17. S. A. Murphy, Ann. Statist. 22, 712 (1994). 18. S. A. Murphy, Ann. Statist. 23, 182 (1995). 19. S. A. Murphy and A. W. van der Vaart, J. Am. Statist. Assoc. 95, 449 (2000). 20. D. Oakes, J. R. Statist. Soc. B 44, 414 (1982). 21. D. Oakes, In Modern Statistical Methods in Chronic Disease Epidemiology, eds. R. L. Prentice and S.H. Moolgavkar (John Wiley, New York, 1986). 22. D. Oakes, J. Am. Statist. Assoc. 84, 487 (1989). 23. D. Oakes and A. K. Manatunga, Biometrika 79, 827 (1992). 24. E. Parner, E. Ann. Statist. 26, 183 (1998). 25. J. Shih and T.A. Louis, Biometrics 5 1 , 1384 (1995).

S U R R O G A T E DATA A N D F R A C T I O N A L B R O W N I A N MOTION PETER RABINOVITCH Alcatel Research and Innovation 600 March Rd. Ottawa ON K2K 2E6, Canada Email: Peter. Rabinovitch@Alcatel, com Statistical analysis of fractional Brownian motion is difficult because it exhibits long range dependence. In this paper, we describe experiments that show that the method of surrogate data can help with this problem.

1

Introduction

Fractional Brownian motion (Mandelbrot and Van Ness 7 ) has become one of the standard models of telecommunications traffic due to three facts. When appropriate, it is a parsimonious model, as it is parameterized by three parameters; it has Gaussian marginal distributions; and it exhibits long range dependence, as many real traffic streams do. The long range dependence, while necessary for accurate modeling, presents statistical difficulties. For example, the standard way of proving the Central Limit Theorem (CLT) for dependent random variables (see Resnick 11 ) is to find an integer m such that random variables further apart than m are independent. Then, in essence, one applies the regular, independent CLT and estimates the error term introduced by leaving out blocks of m random variables, and shows that this error term can be made as small as desired. However, in the long range dependent setting, finding such an m may not be possible. One could also appeal to the generalizations of the bootstrap to dependent data, such as the moving blocks bootstrap. However, it is a known result (Lahiri 6 ) that the moving blocks bootstrap does not necessarily work with long range dependent data. The solution to this problem comes from an unlikely source, nonlinear dynamics. Surrogate data (also known as "phase scrambling", "Fourier bootstrap methods", etc.) was developed for hypothesis testing in nonlinear dynamics in the early 1990's. At that time, many papers were published claiming that some observational set of data had a particular fractional correlation dimension, and therefore that the data was chaotic. The surrogate data method was developed (Theiler et al. 13 ) to be able to test this claim, by generating many linear, Gaussian synthetic data sets that had the same first and second

225

226

order statistics as the original data. One could then compare a statistic, such as correlation dimension, of the original data set to a histogram of the same statistic of the synthetic data sets, and if they were significantly different, one could reject the null hypothesis that the original data could be generated by a linear, Gaussian process. We will present evidence, both by referring to the research literature, and by numerical experiments, that the surrogate data method captures the relevant characteristics of fractional Brownian motion, including long range dependence. Pointer to S+ source code is provided so that the readers can perform their own experiments. 2

Fractional Brownian Motion

Definition 2.1. A stochastic process {X(t),t Brownian motion with Hurst parameter H if

> 0} is said to be a fractional

1. X(t) has stationary increments for t > 0; 2. X(t) is normally distributed with mean 0; 3. X(0) = 0 almost surely; and 4. The increments of X(t),Z(j) Pz(k)

:= X(j + 1) - X(j),j

= ±{\k + l\2H +

= 0,1,... satisfy,

\k-l\2H-2h?h}

where pz denotes the autocorrelation function of Z. Remark 2.1. Let {X(t), t > 0} be a fractional Brownian motion with Hurst parameter H > 1/2. Then the increments of X(t) are long range dependent. There are many ways to simulate fractional Brownian motion, such as random mid-point displacement; superposition of heavy tailed on/off processes, wavelet methods, and Fourier methods. In this paper we use the last method, as described in Paxson 8 . A common model for long range dependent network traffic is W(t) = Xt +

aXH(t),

where W(t) represents the total amount of traffic that has arrived by time t, A and a are parameters representing the mean and standard deviation of the arrivals, and Xn(t) is a fractional Brownian motion with Hurst parameter H.

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3

Surrogate Data

Theiler et al. 13 proposed a method of generating synthetic data that preserves the mean and the autocorrelation of the original series. The mean is preserved by subtracting it from the original series, and then adding it back to the synthetic series. This is standard, as it is frequently also a step in moving blocks bootstrap, or even classical time series analysis [ARMA modeling, etc.). For the autocorrelation, they proposed taking the Fourier transform of the series, and then scrambling the phases of the transformed data, while ensuring that the phases remain symmetric so that when the data is transformed back it is real valued. The phases can be either sampled with replacement from the original phases (i.e. bootstrapped), or they can be sampled uniformly on the interval. Here, we use the latter choice for its simplicity.